Page 1

Econometrics


Badi H. Baltagi

Econometrics

Fourth Edition

123


Professor Badi H. Baltagi Syracuse University Center for Policy Research 426 Eggers Hall Syracuse, NY 13244-1020 USA bbaltagi@maxwell.syr.edu

ISBN 978-3-540-76515-8

e-ISBN 978-3-540-76516-5

DOI 10.1007/978-3-540-76516-5 Library of Congress Control Number: 2007939803 c 2008 Springer-Verlag Berlin Heidelberg  This work is subject to copyright. All rights are reserved, whether the whole or part of the material is concerned, specifically the rights of translation, reprinting, reuse of illustrations, recitation, broadcasting, reproduction on microfilm or in any other way, and storage in data banks. Duplication of this publication or parts thereof is permitted only under the provisions of the German Copyright Law of September 9, 1965, in its current version, and permission for use must always be obtained from Springer. Violations are liable to prosecution under the German Copyright Law. The use of general descriptive names, registered names, trademarks, etc. in this publication does not imply, even in the absence of a specific statement, that such names are exempt from the relevant protective laws and regulations and therefore free for general use. Production: LE-TEX Jelonek, Schmidt & Vรถckler GbR, Leipzig Coverdesign: WMX Design GmbH, Heidelberg Printed on acid-free paper 987654321 springer.com


To My Wife Phyllis


Preface This book is intended for a first year graduate course in econometrics. Courses requiring matrix algebra as a pre-requisite to econometrics can start with Chapter 7. Chapter 2 has a quick refresher on some of the required background needed from statistics for the proper understanding of the material in this book. For an advanced undergraduate/masters class not requiring matrix algebra, one can structure a course based on Chapter 1; Section 2.6 on descriptive statistics; Chapters 3-6; Section 11.1 on simultaneous equations; and Chapter 14 on time-series analysis. This book teaches some of the basic econometric methods and the underlying assumptions behind them. Estimation, hypotheses testing and prediction are three recurrent themes in this book. Some uses of econometric methods include (i) empirical testing of economic theory, whether it is the permanent income consumption theory or purchasing power parity, (ii) forecasting, whether it is GNP or unemployment in the U.S. economy or future sales in the computer industry. (iii) Estimation of price elasticities of demand, or returns to scale in production. More importantly, econometric methods can be used to simulate the effect of policy changes like a tax increase on gasoline consumption, or a ban on advertising on cigarette consumption. It is left to the reader to choose among the available econometric/statistical software to use, like EViews, SAS, STATA, TSP, SHAZAM, Microfit, PcGive, LIMDEP, and RATS, to mention a few. The empirical illustrations in the book utilize a variety of these software packages. Of course, these packages have different advantages and disadvantages. However, for the basic coverage in this book, these differences may be minor and more a matter of what software the reader is familiar or comfortable with. In most cases, I encourage my students to use more than one of these packages and to verify these results using simple programming languages like GAUSS, OX, R and MATLAB. This book is not meant to be encyclopedic. I did not attempt the coverage of Bayesian econometrics simply because it is not my comparative advantage. The reader should consult Koop (2003) for a more recent treatment of the subject. Nonparametrics and semiparametrics are popular methods in today’s econometrics, yet they are not covered in this book to keep the technical difficulty at a low level. These are a must for a follow-up course in econometrics, see Li and Racine (2007). Also, for a more rigorous treatment of asymptotic theory, see White (1984). Despite these limitations, the topics covered in this book are basic and necessary in the training of every economist. In fact, it is but a ‘stepping stone’, a ‘sample of the good stuff’ the reader will find in this young, energetic and ever evolving field. I hope you will share my enthusiasm and optimism in the importance of the tools you will learn when you are through reading this book. Hopefully, it will encourage you to consult the suggested readings on this subject that are referenced at the end of each chapter. In his inaugural lecture at the University of Birmingham, entitled “Econometrics: A View from the Toolroom,” Peter C.B. Phillips (1977) concluded: “the toolroom may lack the glamour of economics as a practical art in government or business, but it is every bit as important. For the tools (econometricians) fashion provide the key to improvements in our quantitative information concerning matters of economic policy.”


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Preface

As a student of econometrics, I have benefited from reading Johnston (1984), Kmenta (1986), Theil (1971), Klein (1974), Maddala (1977), and Judge, et al. (1985), to mention a few. As a teacher of undergraduate econometrics, I have learned from Kelejian and Oates (1989), Wallace and Silver (1988), Maddala (1992), Kennedy (1992), Wooldridge (2003) and Stock and Watson (2003). As a teacher of graduate econometrics courses, Greene (1993), Judge, et al. (1985), Fomby, Hill and Johnson (1984) and Davidson and MacKinnon (1993) have been my regular companions. The influence of these books will be evident in the pages that follow. At the end of each chapter I direct the reader to some of the classic references as well as further suggested readings. This book strikes a balance between a rigorous approach that proves theorems and a completely empirical approach where no theorems are proved. Some of the strengths of this book lie in presenting some difficult material in a simple, yet rigorous manner. For example, Chapter 12 on pooling time-series of cross-section data is drawn from the author’s area of expertise in econometrics and the intent here is to make this material more accessible to the general readership of econometrics. The exercises contain theoretical problems that should supplement the understanding of the material in each chapter. Some of these exercises are drawn from the Problems and Solutions series of Econometric Theory (reprinted with permission of Cambridge University Press). In addition, the book has a set of empirical illustrations demonstrating some of the basic results learned in each chapter. Data sets from published articles are provided for the empirical exercises. These exercises are solved using several econometric software packages and are available in the Solution Manual. This book is by no means an applied econometrics text, and the reader should consult Berndt’s (1991) textbook for an excellent treatment of this subject. Instructors and students are encouraged to get other data sets from the internet or journals that provide backup data sets to published articles. The Journal of Applied Econometrics and the Journal of Business and Economic Statistics are two such journals. In fact, the Journal of Applied Econometrics has a replication section for which I am serving as an editor. In my econometrics course, I require my students to replicate an empirical paper. Many students find this experience rewarding in terms of giving them hands on application of econometric methods that prepare them for doing their own empirical work. I would like to thank my teachers Lawrence R. Klein, Roberto S. Mariano and Robert Shiller who introduced me to this field; James M. Griffin who provided some data sets, empirical exercises and helpful comments, and many colleagues who had direct and indirect influence on the contents of this book including G.S. Maddala, Jan Kmenta, Peter Schmidt, Cheng Hsiao, Tom Wansbeek, Walter Kr¨ amer, Maxwell King, Peter C.B. Phillips, Alberto Holly, Essie Maasoumi, Aris Spanos, Farshid Vahid, Heather Anderson, Arnold Zellner and Bryan Brown. Also, I would like to thank my students Wei-Wen Xiong, Ming-Jang Weng, Kiseok Nam, Dong Li and Gustavo Sanchez who read parts of this book and solved several of the exercises. Werner M¨ uller and Martina Bihn at Springer for their prompt and professional editorial help. I have also benefited from my visits to the University of Arizona, University of California San-Diego, Monash University, the University of Zurich, the Institute of Advanced Studies in Vienna, and the University of Dortmund, Germany. A special thanks to my wife Phyllis whose help and support were essential to completing this book.


Preface

IX

References Berndt, E.R. (1991), The Practice of Econometrics: Classic and Contemporary (Addison-Wesley: Reading, MA). Davidson, R. and J.G. MacKinnon (1993), Estimation and Inference In Econometrics (Oxford University Press: Oxford, MA). Fomby, T.B., R.C. Hill and S.R. Johnson (1984), Advanced Econometric Methods (Springer-Verlag: New York). Greene, W.H. (1993), Econometric Analysis (Macmillan: New York ). Johnston, J. (1984), Econometric Methods , 3rd. Ed., (McGraw-Hill: New York). Judge, G.G., W.E. Griffiths, R.C. Hill, H. L¨ utkepohl and T.C. Lee (1985), The Theory and Practice of Econometrics , 2nd Ed., (John Wiley: New York). Kelejian, H. and W. Oates (1989), Introduction to Econometrics: Principles and Applications , 2nd Ed., (Harper and Row: New York). Kennedy, P. (1992), A Guide to Econometrics (The MIT Press: Cambridge, MA). Klein, L.R. (1974), A Textbook of Econometrics (Prentice-Hall: New Jersey). Kmenta, J. (1986), Elements of Econometrics , 2nd Ed., (Macmillan: New York). Koop, G. (2003), Bayesian Econometrics, (Wiley: New York). Li, Q. and J.S. Racine (2007), Nonparametric Econometrics, (Princeton University Press: New Jersey). Maddala, G.S. (1977), Econometrics (McGraw-Hill: New York). Maddala, G.S. (1992), Introduction to Econometrics (Macmillan: New York). Phillips, P.C.B. (1977), “Econometrics: A View From the Toolroom,” Inaugural Lecture, University of Birmingham, Birmingham, England. Stock, J.H. and M.W. Watson (2003), Introduction to Econometrics , (Addison-Wesley: New York). Theil, H. (1971), Principles of Econometrics (John Wiley: New York). Wallace, T.D. and L. Silver (1988), Econometrics: An Introduction (Addison-Wesley: New York). White, H. (1984), Asymptotic Theory for Econometrics (Academic Press: Florida). Wooldridge, J.M. (2003), Introductory Econometrics , (South-Western: Ohio).

Data The data sets used in this text can be downloaded from the Springer website in Germany. The address is: http://www.springer.com/978-3-540-76515-8. Please select the link “Samples & Supplements” from the right-hand column.


Table of Contents Preface

VII

Table of Contents

XI

Part I

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1 What Is Econometrics? 1.1 Introduction . . . . . . . 1.2 A Brief History . . . . . 1.3 Critiques of Econometrics 1.4 Looking Ahead . . . . . . Notes . . . . . . . . . . . . . . References . . . . . . . . . . . .

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2 Basic Statistical Concepts 2.1 Introduction . . . . . . . 2.2 Methods of Estimation . 2.3 Properties of Estimators 2.4 Hypothesis Testing . . . 2.5 ConďŹ dence Intervals . . . 2.6 Descriptive Statistics . . Notes . . . . . . . . . . . . . . Problems . . . . . . . . . . . . References . . . . . . . . . . . . Appendix . . . . . . . . . . . .

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. . . . . . . . Assumptions . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .

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3 Simple Linear Regression 3.1 Introduction . . . . . . . . . . . . . . . . . 3.2 Least Squares Estimation and the Classical 3.3 Statistical Properties of Least Squares . . . 3.4 Estimation of Ďƒ 2 . . . . . . . . . . . . . . . 3.5 Maximum Likelihood Estimation . . . . . . 3.6 A Measure of Fit . . . . . . . . . . . . . . 3.7 Prediction . . . . . . . . . . . . . . . . . . 3.8 Residual Analysis . . . . . . . . . . . . . . 3.9 Numerical Example . . . . . . . . . . . . . 3.10 Empirical Example . . . . . . . . . . . . . Problems . . . . . . . . . . . . . . . . . . . . . . References . . . . . . . . . . . . . . . . . . . . . . Appendix . . . . . . . . . . . . . . . . . . . . . .

4 Multiple Regression Analysis 73 4.1 Introduction . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 73


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4.2 Least Squares Estimation . . . . . . . . . . . . . . . . . . . . . . . . 4.3 Residual Interpretation of Multiple Regression Estimates . . . . . . 4.4 Overspecification and Underspecification of the Regression Equation 4.5 R-Squared versus R-Bar-Squared . . . . . . . . . . . . . . . . . . . 4.6 Testing Linear Restrictions . . . . . . . . . . . . . . . . . . . . . . . 4.7 Dummy Variables . . . . . . . . . . . . . . . . . . . . . . . . . . . . Note . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . Problems . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . References . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . Appendix . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .

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6 Distributed Lags and Dynamic Models 6.1 Introduction . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 6.2 Infinite Distributed Lag . . . . . . . . . . . . . . . . . . . . . . . . . . . 6.2.1 Adaptive Expectations Model (AEM) . . . . . . . . . . . . . . . 6.2.2 Partial Adjustment Model (PAM) . . . . . . . . . . . . . . . . . 6.3 Estimation and Testing of Dynamic Models with Serial Correlation . . 6.3.1 A Lagged Dependent Variable Model with AR(1) Disturbances 6.3.2 A Lagged Dependent Variable Model with MA(1) Disturbances 6.4 Autoregressive Distributed Lag . . . . . . . . . . . . . . . . . . . . . . . Note . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . Problems . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . References . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .

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5 Violations of the Classical Assumptions 5.1 Introduction . . . . . . . . . . . . . . . 5.2 The Zero Mean Assumption . . . . . . 5.3 Stochastic Explanatory Variables . . . 5.4 Normality of the Disturbances . . . . . 5.5 Heteroskedasticity . . . . . . . . . . . . 5.6 Autocorrelation . . . . . . . . . . . . . Notes . . . . . . . . . . . . . . . . . . . . . . Problems . . . . . . . . . . . . . . . . . . . . References . . . . . . . . . . . . . . . . . . . .

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Part II 7 The 7.1 7.2 7.3 7.4 7.5 7.6 7.7

147 General Linear Model: The Basics Introduction . . . . . . . . . . . . . . . . . . . . . . Least Squares Estimation . . . . . . . . . . . . . . . Partitioned Regression and the Frisch-Waugh-Lovell Maximum Likelihood Estimation . . . . . . . . . . . Prediction . . . . . . . . . . . . . . . . . . . . . . . Confidence Intervals and Test of Hypotheses . . . . Joint Confidence Intervals and Test of Hypotheses .

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Table of Contents

7.8 Restricted MLE and Restricted Least Squares . 7.9 Likelihood Ratio, Wald and Lagrange Multiplier Notes . . . . . . . . . . . . . . . . . . . . . . . . . . . Problems . . . . . . . . . . . . . . . . . . . . . . . . . References . . . . . . . . . . . . . . . . . . . . . . . . . Appendix . . . . . . . . . . . . . . . . . . . . . . . . .

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159 160 165 165 170 171

8 Regression Diagnostics and Specification Tests 8.1 Influential Observations . . . . . . . . . . . . . . . . . . . . 8.2 Recursive Residuals . . . . . . . . . . . . . . . . . . . . . . 8.3 Specification Tests . . . . . . . . . . . . . . . . . . . . . . . 8.4 Nonlinear Least Squares and the Gauss-Newton Regression 8.5 Testing Linear versus Log-Linear Functional Form . . . . . Notes . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . Problems . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . References . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .

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177 177 185 194 204 212 214 214 218

9 Generalized Least Squares 9.1 Introduction . . . . . . . . . . . . . . . . 9.2 Generalized Least Squares . . . . . . . . 9.3 Special Forms of Ω . . . . . . . . . . . . 9.4 Maximum Likelihood Estimation . . . . . 9.5 Test of Hypotheses . . . . . . . . . . . . 9.6 Prediction . . . . . . . . . . . . . . . . . 9.7 Unknown Ω . . . . . . . . . . . . . . . . 9.8 The W, LR and LM Statistics Revisited 9.9 Spatial Error Correlation . . . . . . . . . Note . . . . . . . . . . . . . . . . . . . . . . . . Problems . . . . . . . . . . . . . . . . . . . . . References . . . . . . . . . . . . . . . . . . . . .

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10 Seemingly Unrelated Regressions 10.1 Introduction . . . . . . . . . . . . . . . . . . . . . . . . . . . 10.2 Feasible GLS Estimation . . . . . . . . . . . . . . . . . . . . 10.3 Testing Diagonality of the Variance-Covariance Matrix . . . 10.4 Seemingly Unrelated Regressions with Unequal Observations 10.5 Empirical Example . . . . . . . . . . . . . . . . . . . . . . . Problems . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . References . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .

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11 Simultaneous Equations Model 11.1 Introduction . . . . . . . . . . . . . . . 11.1.1 Simultaneous Bias . . . . . . . . 11.1.2 The Identification Problem . . . 11.2 Single Equation Estimation: Two-Stage 11.2.1 Spatial Lag Dependence . . . .

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11.3 System Estimation: Three-Stage Least Squares 11.4 Test for Over-Identification Restrictions . . . . 11.5 Hausman’s Specification Test . . . . . . . . . . 11.6 Empirical Example: Crime in North Carolina . Notes . . . . . . . . . . . . . . . . . . . . . . . . . . Problems . . . . . . . . . . . . . . . . . . . . . . . . References . . . . . . . . . . . . . . . . . . . . . . . . Appendix . . . . . . . . . . . . . . . . . . . . . . . .

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267 269 271 273 277 277 287 289

12 Pooling Time-Series of Cross-Section Data 12.1 Introduction . . . . . . . . . . . . . . . . . . . . . . . . . . . 12.2 The Error Components Model . . . . . . . . . . . . . . . . . 12.2.1 The Fixed Effects Model . . . . . . . . . . . . . . . . 12.2.2 The Random Effects Model . . . . . . . . . . . . . . 12.2.3 Maximum Likelihood Estimation . . . . . . . . . . . 12.3 Prediction . . . . . . . . . . . . . . . . . . . . . . . . . . . . 12.4 Empirical Example . . . . . . . . . . . . . . . . . . . . . . . 12.5 Testing in a Pooled Model . . . . . . . . . . . . . . . . . . . 12.6 Dynamic Panel Data Models . . . . . . . . . . . . . . . . . . 12.6.1 Empirical Illustration . . . . . . . . . . . . . . . . . . 12.7 Program Evaluation and Difference-in-Differences Estimator 12.7.1 The Difference-in-Differences Estimator . . . . . . . . Problems . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . References . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . .

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323 . 323 . 323 . 324 . 326 . 331 . 332 . 334 . 334 . 335 . 339 . 339 . 340 . 341 . 344 . 345 . 347 . 347 . 351 . 353

13 Limited Dependent Variables 13.1 Introduction . . . . . . . . . . . . . . . . 13.2 The Linear Probability Model . . . . . . 13.3 Functional Form: Logit and Probit . . . . 13.4 Grouped Data . . . . . . . . . . . . . . . 13.5 Individual Data: Probit and Logit . . . . 13.6 The Binary Response Model Regression . 13.7 Asymptotic Variances for Predictions and 13.8 Goodness of Fit Measures . . . . . . . . . 13.9 Empirical Examples . . . . . . . . . . . . 13.10 Multinomial Choice Models . . . . . . . . 13.10.1 Ordered Response Models . . . . 13.10.2 Unordered Response Models . . . 13.11 The Censored Regression Model . . . . . 13.12 The Truncated Regression Model . . . . 13.13 Sample Selectivity . . . . . . . . . . . . . Notes . . . . . . . . . . . . . . . . . . . . . . . Problems . . . . . . . . . . . . . . . . . . . . . References . . . . . . . . . . . . . . . . . . . . . Appendix . . . . . . . . . . . . . . . . . . . . .

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Table of Contents

14 Time-Series Analysis 14.1 Introduction . . . . . . . . . . . . . . . . . . . 14.2 Stationarity . . . . . . . . . . . . . . . . . . . 14.3 The Box and Jenkins Method . . . . . . . . . 14.4 Vector Autoregression . . . . . . . . . . . . . . 14.5 Unit Roots . . . . . . . . . . . . . . . . . . . . 14.6 Trend Stationary versus Difference Stationary 14.7 Cointegration . . . . . . . . . . . . . . . . . . 14.8 Autoregressive Conditional Heteroskedasticity Note . . . . . . . . . . . . . . . . . . . . . . . . . . . Problems . . . . . . . . . . . . . . . . . . . . . . . . References . . . . . . . . . . . . . . . . . . . . . . . .

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355 355 355 356 360 361 365 366 368 371 371 375

Appendix

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List of Figures

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List of Tables

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Index

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Part I


CHAPTER 1

What Is Econometrics? 1.1

Introduction

What is econometrics? A few definitions are given below: The method of econometric research aims, essentially, at a conjunction of economic theory and actual measurements, using the theory and technique of statistical inference as a bridge pier. Trygve Haavelmo (1944)

Econometrics may be defined as the quantitative analysis of actual economic phenomena based on the concurrent development of theory and observation, related by appropriate methods of inference. Samuelson, Koopmans and Stone (1954)

Econometrics is concerned with the systematic study of economic phenomena using observed data. Aris Spanos (1986)

Broadly speaking, econometrics aims to give empirical content to economic relations for testing economic theories, forecasting, decision making, and for ex post decision/policy evaluation. J. Geweke, J. Horowitz, and M.H. Pesaran (2007)

For other definitions of econometrics, see Tintner (1953). An econometrician has to be a competent mathematician and statistician who is an economist by training. Fundamental knowledge of mathematics, statistics and economic theory are a necessary prerequisite for this field. As Ragnar Frisch (1933) explains in the first issue of Econometrica, it is the unification of statistics, economic theory and mathematics that constitutes econometrics. Each view point, by itself is necessary but not sufficient for a real understanding of quantitative relations in modern economic life. Ragnar Frisch is credited with coining the term ‘econometrics’ and he is one of the founders of the Econometrics Society, see Christ (1983). Econometrics aims at giving empirical content to economic relationships. The three key ingredients are economic theory, economic data, and statistical methods. Neither ‘theory without measurement’, nor ‘measurement without theory’ are sufficient for explaining economic phenomena. It is as Frisch emphasized their union that is the key for success in the future development of econometrics. Lawrence R. Klein, the 1980 recipient of the Nobel Prize in economics “for the creation of econometric models and their application to the analysis of economic fluctuations and economic policies,”1 has always emphasized the integration of economic theory, statistical methods and practical economics. The exciting thing about econometrics is its concern for verifying or refuting economic laws, such as purchasing power parity, the life cycle hypothesis, the quantity theory of money, etc. These economic laws or hypotheses are testable with economic data. In fact, David F. Hendry (1980) emphasized this function of econometrics:


4

Chapter 1: What Is Econometrics?

The three golden rules of econometrics are test, test and test; that all three rules are broken regularly in empirical applications is fortunately easily remedied. Rigorously tested models, which adequately described the available data, encompassed previous findings and were derived from well based theories would enhance any claim to be scientific. Econometrics also provides quantitative estimates of price and income elasticities of demand, estimates of returns to scale in production, technical efficiency, the velocity of money, etc. It also provides predictions about future interest rates, unemployment, or GNP growth. Lawrence Klein (1971) emphasized this last function of econometrics: Econometrics had its origin in the recognition of empirical regularities and the systematic attempt to generalize these regularities into “laws” of economics. In a broad sense, the use of such “laws” is to make predictions - - about what might have or what will come to pass. Econometrics should give a base for economic prediction beyond experience if it is to be useful. In this broad sense it may be called the science of economic prediction. Econometrics, while based on scientific principles, still retains a certain element of art. According to Malinvaud (1966), the art in econometrics is trying to find the right set of assumptions which are sufficiently specific, yet realistic to enable us to take the best possible advantage of the available data. Data in economics are not generated under ideal experimental conditions as in a physics laboratory. This data cannot be replicated and is most likely measured with error. In some cases, the available data are proxies for variables that are either not observed or cannot be measured. Many published empirical studies find that economic data may not have enough variation to discriminate between two competing economic theories. Manski (1995, p. 8) argues that Social scientists and policymakers alike seem driven to draw sharp conclusions, even when these can be generated only by imposing much stronger assumptions than can be defended. We need to develop a greater tolerance for ambiguity. We must face up to the fact that we cannot answer all of the questions that we ask. To some, the “art” element in econometrics has left a number of distinguished economists doubtful of the power of econometrics to yield sharp predictions. In his presidential address to the American Economic Association, Wassily Leontief (1971, pp. 2-3) characterized econometrics work as: an attempt to compensate for the glaring weakness of the data base available to us by the widest possible use of more and more sophisticated techniques. Alongside the mounting pile of elaborate theoretical models we see a fast growing stock of equally intricate statistical tools. These are intended to stretch to the limit the meager supply of facts. Most of the time the data collected are not ideal for the economic question at hand because they were posed to answer legal requirements or comply to regulatory agencies. Griliches (1986, p.1466) describes the situation as follows:


1.2

A Brief History

5

Econometricians have an ambivilant attitude towards economic data. At one level, the ‘data’ are the world that we want to explain, the basic facts that economists purport to elucidate. At the other level, they are the source of all our trouble. Their imperfections make our job difficult and often impossible... We tend to forget that these imperfections are what gives us our legitimacy in the first place... Given that it is the ‘badness’ of the data that provides us with our living, perhaps it is not all that surprising that we have shown little interest in improving it, in getting involved in the grubby task of designing and collecting original data sets of our own. Most of our work is on ‘found’ data, data that have been collected by somebody else, often for quite different purposes. Even though economists are increasingly getting involved in collecting their data and measuring variables more accurately and despite the increase in data sets and data storage and computational accuracy, some of the warnings given by Griliches (1986, p. 1468) are still valid today: The encounters between econometricians and data are frustrating and ultimately unsatisfactory both because econometricians want too much from the data and hence tend to be dissappointed by the answers, and because the data are incomplete and imperfect. In part it is our fault, the appetite grows with eating. As we get larger samples, we keep adding variables and expanding our models, until on the margin, we come back to the same insignificance levels.

1.2

A Brief History

For a brief review of the origins of econometrics before World War II and its development in the 1940-1970 period, see Klein (1971). Klein gives an interesting account of the pioneering works of Moore (1914) on economic cycles, Working (1927) on demand curves, Cobb and Douglas (1928) on the theory of production, Schultz (1938) on the theory and measurement of demand, and Tinbergen (1939) on business cycles. As Klein (1971, p. 415) adds: The works of these men mark the beginnings of formal econometrics. Their analysis was systematic, based on the joint foundations of statistical and economic theory, and they were aiming at meaningful substantive goals - to measure demand elasticity, marginal productivity and the degree of macroeconomic stability. The story of the early progress in estimating economic relationships in the U.S. is given in Christ (1985). The modern era of econometrics, as we know it today, started in the 1940’s. Klein (1971) attributes the formulation of the econometrics problem in terms of the theory of statistical inference to Haavelmo (1943, 1944) and Mann and Wald (1943). This work was extended later by T.C. Koopmans, J. Marschak, L. Hurwicz, T.W. Anderson and others at the Cowles Commission in the late 1940’s and early 1950’s, see Koopmans (1950). Klein (1971, p. 416) adds: At this time econometrics and mathematical economics had to fight for academic recognition. In retrospect, it is evident that they were growing disciplines and becoming increasingly attractive to the new generation of economic students after World War II, but only a few of the largest and most advanced universities offered formal work in these subjects. The mathematization of economics was strongly resisted.


6

Chapter 1: What Is Econometrics?

This resistance is a thing of the past, with econometrics being an integral part of economics, taught and practiced worldwide. Econometrica, the official journal of the Econometric Society is one of the leading journals in economics, and today the Econometric Society boast a large membership worldwide. Today, it is hard to read any professional article in leading economics and econometrics journals without seeing mathematical equations. Students of economics and econometrics have to be proficient in mathematics to comprehend this research. In an Econometric Theory interview, professor J. D. Sargan of the London School of Economics looks back at his own career in econometrics and makes the following observations: “... econometric theorists have really got to be much more professional statistical theorists than they had to be when I started out in econometrics in 1948... Of course this means that the starting econometrician hoping to do a Ph.D. in this field is also finding it more difficult to digest the literature as a prerequisite for his own study, and perhaps we need to attract students of an increasing degree of mathematical and statistical sophistication into our field as time goes by,” see Phillips (1985, pp. 134-135). This is also echoed by another giant in the field, professor T.W. Anderson of Stanford, who said in an Econometric Theory interview: “These days econometricians are very highly trained in mathematics and statistics; much more so than statisticians are trained in economics; and I think that there will be more cross-fertilization, more joint activity,” see Phillips (1986, p. 280). Research at the Cowles Commission was responsible for providing formal solutions to the problems of identification and estimation of the simultaneous equations model, see Christ (1985).2 Two important monographs summarizing much of the work of the Cowles Commission at Chicago, are Koopmans and Marschak (1950) and Koopmans and Hood (1953).3 The creation of large data banks of economic statistics, advances in computing, and the general acceptance of Keynesian theory, were responsible for a great flurry of activity in econometrics. Macroeconometric modelling started to flourish beyond the pioneering macro models of Klein (1950) and Klein and Goldberger (1955). For the story of the founding of Econometrica and the Econometric Society, see Christ (1983). Suggested readings on the history of econometrics are Pesaran (1987), Epstein (1987) and Morgan (1990). In the conclusion of her book on The History of Econometric Ideas, Morgan (1990; p. 264) explains: In the first half of the twentieth century, econometricians found themselves carrying out a wide range of tasks: from the precise mathematical formulation of economic theories to the development tasks needed to build an econometric model; from the application of statistical methods in data preperation to the measurement and testing of models. Of necessity, econometricians were deeply involved in the creative development of both mathematical economic theory and statistical theory and techniques. Between the 1920s and the 1940s, the tools of mathematics and statistics were indeed used in a productive and complementary union to forge the essential ideas of the econometric approach. But the changing nature of the econometric enterprise in the 1940s caused a return to the division of labour favoured in the late nineteenth century, with mathematical economists working on theory building and econometricians concerned with statistical work. By the 1950s the founding ideal of econometrics, the union of mathematical and statistical economics into a truly synthetic economics, had collapsed.


1.3

Critiques of Econometrics

7

In modern day usage, econometrics have become the application of statistical methods to economics, like biometrics and psychometrics. Although, the ideals of Frisch still live on in Econometrica and the Econometric Society, Maddala (1999) argues that: “In recent years the issues of Econometrica have had only a couple of papers in econometrics (statistical methods in economics) and the rest are all on game theory and mathematical economics. If you look at the list of fellows of the Econometric Society, you find one or two econometricians and the rest are game theorists and mathematical economists.” This may be a little exagerated but it does summarize the rift between modern day econometrics and mathematical economics. For a recent world wide ranking of econometricians as well as academic institutions in the field of econometrics, see Baltagi (2007).

1.3

Critiques of Econometrics

Econometrics has its critics. Interestingly, John Maynard Keynes (1940, p. 156) had the following to say about Jan Tinbergen’s (1939) pioneering work: No one could be more frank, more painstaking, more free of subjective bias or parti pris than Professor Tinbergen. There is no one, therefore, so far as human qualities go, whom it would be safer to trust with black magic. That there is anyone I would trust with it at the present stage or that this brand of statistical alchemy is ripe to become a branch of science, I am not yet persuaded. But Newton, Boyle and Locke all played with alchemy. So let him continue.4 In 1969, Jan Tinbergen shared the first Nobel Prize in economics with Ragnar Frisch. Recent well cited critiques of econometrics include the Lucas (1976) critique which is based on the Rational Expectations Hypothesis (REH). As Pesaran (1990, p. 17) puts it: The message of the REH for econometrics was clear. By postulating that economic agents form their expectations endogenously on the basis of the true model of the economy and a correct understanding of the processes generating exogenous variables of the model, including government policy, the REH raised serious doubts about the invariance of the structural parameters of the mainstream macroeconometric models in face of changes in government policy. Responses to this critique include Pesaran (1987). Other lively debates among econometricians include Ed Leamer’s (1983) article entitled “Let’s Take the Con Out of Econometrics,” and the response by McAleer, Pagan and Volker (1985). Rather than leave the reader with criticisms of econometrics especially before we embark on the journey to learn the tools of the trade, we conclude this section with the following quote from Pesaran (1990, pp. 25-26): There is no doubt that econometrics is subject to important limitations, which stem largely from the incompleteness of the economic theory and the non-experimental nature of economic data. But these limitations should not distract us from recognizing the fundamental role that econometrics has come to play in the development of economics as a scientific discipline. It may not be possible conclusively to reject economic theories by means of econometric methods, but it does not mean that


8

Chapter 1: What Is Econometrics?

nothing useful can be learned from attempts at testing particular formulations of a given theory against (possible) rival alternatives. Similarly, the fact that econometric modelling is inevitably subject to the problem of specification searches does not mean that the whole activity is pointless. Econometric models are important tools for forecasting and policy analysis, and it is unlikely that they will be discarded in the future. The challenge is to recognize their limitations and to work towards turning them into more reliable and effective tools. There seem to be no viable alternatives.

1.4

Looking Ahead

Econometrics have experienced phenomenal growth in the past 50 years. There are five volumes of the Handbook of Econometrics running to 3833 pages. Most of it dealing with post 1960’s research. A lot of the recent growth reflects the rapid advances in computing technology. The broad availability of micro data bases is a major advance which facilitated the growth of panel data methods (see Chapter 12) and microeconometric methods especially on sample selection and discrete choice (see Chapter 13) and that also lead to the award of the Nobel Prize in Economics to James Heckman and Daniel McFadden in 2000. The explosion in research in time series econometrics which lead to the development of ARCH and GARCH and cointegration (see Chapter 14) which also lead to the award of the Nobel Prize in Economics to Clive Granger and Robert Engle in 2003. It is a different world than it was 30 years ago. The computing facilities changed dramatically. The increasing accessibility of cheap and powerful computing facilities are helping to make the latest econometric methods more readily available to applied researchers. Today, there is hardly a field in economics which has not been intensive in its use of econometrics in empirical work. Pagan (1987, p. 81) observed that the work of econometric theorists over the period 1966-1986 have become part of the process of economic investigation and the training of economists. Based on this criterion, he declares econometrics as an “outstanding success.” He adds that: The judging of achievement inevitably involves contrast and comparison. Over a period of twenty years this would be best done by interviewing a time-travelling economist displaced from 1966 to 1986. I came into econometrics just after the beginning of this period, so have some appreciation for what has occurred. But because I have seen the events gradually unfolding, the effects upon me are not as dramatic. Nevertheless, let me try to be a time-traveller and comment on the perceptions of a 1966’er landing in 1986. My first impression must be of the large number of people who have enough econometric and computer skills to formulate, estimate and simulate highly complex and non-linear models. Someone who could do the equivalent tasks in 1966 was well on the way to a Chair. My next impression would be of the widespread use and purchase of econometric services in the academic, government, and private sectors. Quantification is now the norm rather than the exception. A third impression, gleaned from a sounding of the job market, would be a persistent tendency towards an excess demand for well-trained econometricians. The economist in me would have to acknowledge that the market judges the products of the discipline as a success.


Notes

9

The challenge for the 21st century is to narrow the gap between theory and practice. Many feel that this gap has been widening with theoretical research growing more and more abstract and highly mathematical without an application in sight or a motivation for practical use. Heckman (2001) argues that econometrics is useful only if it helps economists conduct and interpret empirical research on economic data. He warns that the gap between econometric theory and empirical practice has grown over the past two decades. Theoretical econometrics becoming more closely tied to mathematical statistics. Although he finds nothing wrong, and much potential value, in using methods and ideas from other fields to improve empirical work in economics, he does warn of the risks involved in uncritically adopting the methods and mind set of the statisticians: Econometric methods uncritically adapted from statistics are not useful in many research activities pursued by economists. A theorem-proof format is poorly suited for analyzing economic data, which requires skills of synthesis, interpretation and empirical investigation. Command of statistical methods is only a part, and sometimes a very small part, of what is required to do first class empirical research. In an Econometric Theory interview with Jan Tinbergen, Magnus and Morgan (1987, p.117) describe Tinbergen as one of the founding fathers of econometrics, publishing in the field from 1927 until the early 1950s. They add: “Tinbergen’s approach to economics has always been a practical one. This was highly appropriate for the new field of econometrics, and enabled him to make important contributions to conceptual and theoretical issues, but always in the context of a relevant economic problem.” The founding fathers of econometrics have always had the practitioner in sight. This is a far cry from many theoretical econometricians who refrain from applied work. The recent entry by Geweke, Horowitz, and Pesaran (2007) in the The New Palgrave Dictionary provides the following recommendations for the future: Econometric theory and practice seek to provide information required for informed decision-making in public and private economic policy. This process is limited not only by the adequacy of econometrics, but also by the development of economic theory and the adequacy of data and other information. Effective progress, in the future as in the past, will come from simultaneous improvements in econometrics, economic theory, and data. Research that specifically addresses the effectiveness of the interface between any two of these three in improving policy — to say nothing of all of them — necessarily transcends traditional subdisciplinary boundaries within economics. But it is precisely these combinations that hold the greatest promise for the social contribution of academic economics.

Notes 1. See the interview of Professor L.R. Klein by Mariano (1987). Econometric Theory publishes interviews with some of the giants in the field. These interviews offer a wonderful glimpse at the life and work of these giants. 2. Simultaneous equations model is an integral part of econometrics and is studied in Chapter 11.


10

Chapter 1: What Is Econometrics? 3. Tjalling Koopmans was the joint recipient of the Nobel Prize in Economics in 1975. In addition to his work on the identification and estimation of simultaneous equations models, he received the Nobel Prize for his work in optimization and economic theory. 4. I encountered this attack by Keynes on Tinbergen in the inaugural lecture that Peter C.B. Phillips (1977) gave at the University of Birmingham entitled “Econometrics: A View From the Toolroom,” and David F. Hendry’s (1980) article entitled “Econometrics - Alchemy or Science?”

References Baltagi, B.H. (2007), “Worldwide Econometrics Rankings: 1989-2005,” Econometric Theory, 23: 9521012. Christ, C.F. (1983), “The Founding of the Econometric Society and Econometrica,” Econometrica, 51: 3-6. Christ, C.F. (1985), “Early Progress in Estimating Quantiative Economic Relations in America,” American Economic Review, 12: 39-52. Cobb, C.W. and P.H. Douglas (1928), “A Theory of Production,” American Economic Review, Supplement 18: 139-165. Epstein, R.J. (1987), A History of Econometrics (North-Holland: Amsterdam). Frisch, R. (1933), “Editorial,” Econometrica, 1: 1-14. Geweke, J., J. Horowitz, and M. H.Pesaran (2007), “Econometrics: A Bird’s Eye View,” forthcoming in The New Palgrave Dictionary, Second Edition. Griliches, Z. (1986), “Economic Data Issues,” in Z. Griliches and M.D. Intriligator (eds), Handbook of Econometrics Vol. III (North Holland: Amsterdam). Haavelmo, T. (1943), “The Statistical Implications of a System of Simultaneous Equations,” Econometrica, 11: 1-12. Haavelmo, T. (1944), “The Probability Approach in Econometrics,” Econometrica, Supplement to Volume 12: 1-118. Heckman, J.J. (2001), “Econometrics and Empirical Economics,” Journal of Econometrics, 100: 3-5. Hendry, D.F. (1980), “Econometrics - Alchemy or Science?” Economica, 47: 387-406. Keynes, J.M. (1940), “On Method of Statistical Research: Comment,” Economic Journal, 50: 154-156. Klein, L.R. (1971), “Whither Econometrics?” Journal of the American Statistical Association, 66: 415421. Klein, L.R. (1950), Economic Fluctuations in the United States 1921-1941, Cowles Commission Monograph, No. 11 (John Wiley: New York). Klein, L.R. and A.S. Goldberger (1955), An Econometric Model of the United States 1929-1952 (NorthHolland: Amsterdam). Koopmans, T.C. (1950), ed., Statistical Inference in Dynamic Economic Models (John Wiley: New York). Koopmans, T.C. and W.C. Hood (1953), Studies in Econometric Method (John Wiley: New York). Koopmans, T.C. and J. Marschak (1950), eds., Statistical Inference in Dynamic Economic Models (John Wiley: New York).


References

11

Leamer, E.E. (1983), “Lets Take the Con Out of Econometrics,” American Economic Review, 73: 31-43. Leontief, W. (1971), “Theoretical Assumptions and Nonobserved Facts,” American Economic Review, 61: 1-7. Lucas, R.E. (1976), “Econometric Policy Evaluation: A Critique,” in K. Brunner and A.M. Meltzer, eds., The Phillips Curve and Labor Markets, Carnegie Rochester Conferences on Public Policy, 1: 19-46. Maddala, G.S. (1999), “Econometrics in the 21st Century,” in C.R. Rao and R. Szekeley, eds., Statistics for the 21 st Century (Marcel Dekker: New York). Magnus , J.R. and M.S. Morgan (1987), “The ET Interview: Professor J. Tinbergen,” Econometric Theory, 3: 117-142. Malinvaud, E. (1966), Statistical Methods of Econometrics (North-Holland: Amsterdam). Manski, C.F. (1995), Identification Problems in the Social Sciences (Harvard University Press: Cambridge). Mann, H.B. and A. Wald (1943), “On the Statistical Treatment of Linear Stochastic Difference Equations,” Econometrica, 11: 173-220. Mariano, R.S. (1987), “The ET Interview: Professor L.R. Klein,” Econometric Theory, 3: 409-460. McAleer, M., A.R. Pagan and P.A. Volker (1985), “What Will Take The Con Out of Econometrics,” American Economic Review, 75: 293-307. Moore, H.L. (1914), Economic Cycles: Their Law and Cause (Macmillan: New York). Morgan, M. (1990), The History of Econometric Ideas (Cambridge University Press: Cambridge, MA). Pagan, A. (1987), “Twenty Years After: Econometrics, 1966-1986,” paper presented at CORE’s 20th Anniversary Conference, Louvain-la-Neuve. Pesaran, M.H. (1987), The Limits to Rational Expectations (Basil Blackwell: Oxford, MA). Pesaran, M.H. (1990), “Econometrics,” in J. Eatwell, M. Milgate and P. Newman; The New Palgrave: Econometrics (W.W. Norton and Company: New York). Phillips, P.C.B. (1977), “Econometrics: A View From the Toolroom,” Inaugural Lecture, University of Birmingham, Birmingham, England. Phillips, P.C.B. (1985), “ET Interviews: Professor J. D. Sargan,” Econometric Theory, 1: 119-139. Phillips, P.C.B. (1986), “The ET Interview: Professor T. W. Anderson,” Econometric Theory, 2: 249-288. Samuelson, P.A., T.C. Koopmans and J.R.N. Stone (1954), “Report of the Evaluative Committee for Econometrica,” Econometrica, 22: 141-146. Schultz, H. (1938), The Theory and Measurement of Demand (University of Chicago Press: Chicago, IL). Spanos, A. (1986), Statistical Foundations of Econometric Modelling (Cambridge University Press: Cambridge, MA). Tinbergen, J. (1939), Statistical Testing of Business Cycle Theories, Vol. II: Business Cycles in the USA, 1919-1932 (League of Nations: Geneva). Tintner, G. (1953), “The Definition of Econometrics,” Econometrica, 21: 31-40. Working, E.J. (1927), “What Do Statistical ‘Demand Curves’ Show?” Quarterly Journal of Economics, 41: 212-235.


CHAPTER 2

Basic Statistical Concepts 2.1

Introduction

One chapter cannot possibly review what one learned in one or two pre-requisite courses in statistics. This is an econometrics book, and it is imperative that the student have taken at least one solid course in statistics. The concepts of a random variable, whether discrete or continuous, and the associated probability function or probability density function (p.d.f.) are assumed known. Similarly, the reader should know the following statistical terms: Cumulative distribution function, marginal, conditional and joint p.d.f.’s. The reader should be comfortable with computing mathematical expectations, and familiar with the concepts of independence, Bayes Theorem and several continuous and discrete probability distributions. These distributions include: the Bernoulli, Binomial, Poisson, Geometric, Uniform, Normal, Gamma, Chi-squared (χ2 ), Exponential, Beta, t and F distributions. Section 2.2 reviews two methods of estimation, while section 2.3 reviews the properties of the resulting estimators. Section 2.4 gives a brief review of test of hypotheses, while section 2.5 discusses the meaning of confidence intervals. These sections are fundamental background for this book, and the reader should make sure that he or she is familiar with these concepts. Also, be sure to solve the exercises at the end of this chapter.

2.2

Methods of Estimation

Consider a Normal distribution with mean μ and variance σ 2 . This is the important “Gaussian” distribution which is symmetric and bell-shaped and completely determined by its measure of centrality, its mean μ and its measure of dispersion, its variance σ 2 . μ and σ 2 are called the population parameters. Draw a random sample X1 , . . . , Xn independent and identically ¯ and σ 2 by distributed (IID) from this population. We usually estimate μ by μ =X s2 =

n

i=1 (Xi

¯ 2 /(n − 1). − X)

¯ = sample average of incomes of For example, μ = mean income of a household in Houston. X 100 households randomly interviewed in Houston. This estimator of μ could have been obtained by either of the following two methods of estimation:

(i) Method of Moments Simply stated, this method of estimation uses the following rule: Keep equating population moments to their sample counterpart until you have estimated all the population parameters.


14

Chapter 2: Basic Statistical Concepts

Sample

Population

n ¯ Xi /n = X i=1 n 2 i=1 Xi /n .. . n r i=1 Xi /n

E(X) = μ E(X 2 ) = μ2 + σ 2 .. . E(X r )

The normal density is completely identified by μ and σ 2 , hence only the first 2 equations are needed  ¯ μ =X and μ 2 + σ 2 = ni=1 Xi2 /n Substituting the first equation in the second one obtains  ¯ 2 = n (Xi − X) ¯ 2 /n σ 2 = ni=1 Xi2 /n − X i=1

(ii) Maximum Likelihood Estimation (MLE)

For a random sample of size n from the Normal distribution Xi ∼ N (μ, σ 2 ), we have √   − ∞ < Xi < +∞ fi (Xi ; μ, σ 2 ) = (1/σ 2π) exp −(Xi − μ)2 /2σ 2

Since X1 , . . . , Xn are independent and identically distributed, the joint probability density function is given as the product of the marginal probability density functions: n     fi (Xi ; μ, σ 2 ) = (1/2πσ 2 )n/2 exp − ni=1 (Xi − μ)2 /2σ 2 (2.1) f (X1 , . . . , Xn ; μ, σ 2 ) = i=1

Usually, we observe only one sample of n households which could have been generated by any pair of (μ, σ 2 ) with −∞ < μ < +∞ and σ 2 > 0. For each pair, say (μ0 , σ 20 ), f (X1 , . . . , Xn ; μ0 , σ 20 ) denotes the probability (or likelihood) of obtaining that sample. By varying (μ, σ 2 ) we get different probabilities of obtaining this sample. Intuitively, we choose the values of μ and σ 2 that maximize the probability of obtaining this sample. Mathematically, we treat f (X1 , . . . , Xn ; μ, σ 2 ) as L(μ, σ 2 ) and we call it the likelihood function. Maximizing L(μ, σ 2 ) with respect to μ and σ 2 , one gets the first-order conditions of maximization: (∂L/∂μ) = 0

(∂L/∂σ 2 ) = 0

and

Equivalently, we can maximize logL(μ, σ 2 ) rather than L(μ, σ 2 ) and still get the same answer. Usually, the latter monotonic transformation of the likelihood is easier to maximize and the first-order conditions become (∂logL/∂μ) = 0

and

(∂logL/∂σ 2 ) = 0

For the Normal distribution example, we get logL(μ; σ 2 ) = −(n/2)log σ 2 − (n/2)log 2π − (1/2σ 2 ) ∂logL(μ; σ 2 )/∂μ = (1/σ 2 )

n

i=1 (Xi

n

i=1 (Xi

i=1 (Xi

¯ − μ) = 0 ⇒ μ M LE = X

∂logL(μ; σ 2 )/∂σ 2 = −(n/2)(1/σ 2 ) + ⇒σ 2M LE =

n

−μ M LE )2 /n =

n

i=1 (Xi

n

i=1 (Xi

− μ)2 /2σ 4 = 0 ¯ 2 /n − X)

− μ)2


2.2

Methods of Estimation

15

Note that the moments estimators and the maximum likelihood estimators are the same for the Normal distribution example. In general, the two methods need not necessarily give the same estimators. Also, note that the moments estimators will always have the same estimating equations, for example, the first two equations are always E(X) = μ ≡

n

i=1 Xi /n

¯ =X

and

E(X 2 ) = μ2 + σ 2 ≡

n

2 i=1 Xi /n.

For a specific distribution, we need only substitute the relationship between the population moments and the parameters of that distribution. Again, the number of equations needed depends upon the number of parameters of the underlying distribution. For e.g., the exponential distribution has one parameter and needs only one equation whereas the gamma distribution has two parameters and needs two equations. Finally, note that the maximum likelihood technique is heavily reliant on the form of the underlying distribution, but it has desirable properties when it exists. These properties will be discussed in the next section. So far we have dealt with the Normal distribution to illustrate the two methods of estimation. We now apply these methods to the Bernoulli distribution and leave other distributions applications to the exercises. We urge the student to practice on these exercises. Bernoulli Example: In various cases in real life the outcome of an event is binary, a worker may join the labor force or may not. A criminal may return to crime after parole or may not. A television off the assembly line may be defective or not. A coin tossed comes up head or tail, and so on. In this case θ = Pr[Head] and 1 − θ = Pr[Tail] with 0 < θ < 1 and this can be represented by the discrete probability function f (X; θ) = θX (1 − θ)1−X =0

X = 0, 1 elsewhere

The Normal distribution is a continuous distribution since it takes values for all X over the real line. The Bernoulli distribution is discrete, because it is defined only at integer values for X. Note that P [X = 1] = f (1; θ) = θ and P [X = 0] = f (0; θ) = 1 − θ for all values of 0 < θ < 1. A random sample of size n drawn from this distribution will have a joint probability function n n L(θ) = f (X1 , . . . , Xn ; θ) = θ i=1 Xi (1 − θ)n− i=1 Xi with Xi = 0, 1 for i = 1, . . . , n. Therefore,   logL(θ) = ( ni=1 Xi )logθ + (n − ni=1 Xi )log(1 − θ) n  (n − ni=1 Xi ) ∂logL(θ) i=1 Xi = − ∂θ θ (1 − θ)

Solving this first-order condition for θ, one gets   ( ni=1 Xi )(1 − θ) − θ(n − ni=1 Xi ) = 0

which reduces to

  ¯ θM LE = ni=1 Xi /n = X.

This is the frequency of heads in n tosses of a coin.


16

Chapter 2: Basic Statistical Concepts

For the method of moments, we need E(X) =

1

X=0 Xf (X, θ)

= 1.f (1, θ) + 0.f (0, θ) = f (1, θ) = θ

¯ to get  ¯ Once again, the MLE and the method of moments yield and this is equated to X θ = X. the same estimator. Note that only one parameter θ characterizes this Bernoulli distribution and one does not need to equate second or higher population moments to their sample values.

2.3

Properties of Estimators

(i) Unbiasedness μ  is said to be unbiased for μ if and only if E( μ) = μ  ¯ we have E(X) ¯ = n E(Xi )/n = μ and X ¯ is unbiased for μ. No distributional For μ  = X, i=1 assumption is needed as long as the Xi ’s are distributed with the same mean μ. Unbiasedness means that “on the average” our estimator is on target. Let us explain this last statement. If we repeat our drawing of a random sample of 100 households, say 200 times, then we get 200 ¯ Some of these X ¯ ’s will be above μ some below μ, but their average should be very close X’s. to μ. Since in real life situations, we observe only one random sample, there is little consolation ¯ is far from μ. But the larger n is the smaller is the dispersion of this X, ¯ since if our observed X 2 ¯ ¯ var(X) = σ /n and the lesser is the likelihood of this X to be very far from μ. This leads us to the concept of efficiency. (ii) Efficiency For two unbiased estimators, we compare their efficiencies by the ratio of their variances. We say ¯ 2 = X, that the one with lower variance is more efficient. For example, taking μ 1 = X1 versus μ both estimators are unbiased but var( μ1 ) = σ 2 whereas, var( μ2 ) = σ 2 /n and {the relative efficiency of μ 1 with respect to μ 2 } = var( μ2 )/var( μ1 ) = 1/n, see Figure 2.1. To compare all unbiased estimators, we find the one with minimum variance. Such an estimator if it exists is called the MVU (minimum variance unbiased estimator). A lower bound for the variance of any unbiased estimator μ  of μ, is known in the statistical literature as the Cram´er-Rao lower bound, and is given by var( μ) ≥ 1/n{E(∂logf (X; μ))/∂μ}2 = −1/{nE(∂ 2 logf (X; μ))/∂μ2 }

(2.2)

where we use either representation of the bound on the right hand side of (2.2) depending on which one is the simplest to derive. Example 1: Consider the normal density logf (Xi ; μ) = (−1/2)logσ 2 − (1/2)log2π − (1/2)(Xi − μ)2 /σ 2 ∂logf (Xi ; μ)/∂μ = (Xi − μ)/σ 2 ∂ 2 logf (Xi ; μ)/∂μ2 = −(1/σ 2 )


2.3

Properties of Estimators

17

f(x)

X ~ ( m , s 2 / n)

X i ~ ( m, s 2 )

m

x

Figure 2.1 Efficiency Comparisons

with E{∂ 2 logf (Xi ; μ)/∂μ2 } = −(1/σ 2 ). Therefore, the variance of any unbiased estimator of μ, say μ  satisfies the property that var( μ) ≥ σ 2 /n. 2 2 Turning to σ ; let θ = σ , then logf (Xi ; θ) = −(1/2)logθ − (1/2)log2π − (1/2)(Xi − μ)2 /θ

∂logf (Xi ; θ)/∂θ = −1/2θ + (Xi − μ)2 /2θ 2 = {(Xi − μ)2 − θ}/2θ 2 ∂ 2 logf (Xi ; θ)/∂θ 2 = 1/2θ 2 − (Xi − μ)2 /θ 3 = {θ − 2(Xi − μ)2 }/2θ 3 E[∂ 2 logf (Xi ; θ)/∂θ 2 ] = −(1/2θ 2 ), since E(Xi − μ)2 = θ. Hence, for any unbiased estimator of θ, say  θ, its variance satisfies the following property var( θ) ≥ 2θ2 /n, or var( σ 2 ) ≥ 2σ 4 /n. Note that, if one finds an unbiased estimator whose variance attains the Cram´er-Rao lower bound, then this is the MVU estimator. It is important to remember that this is only a lower ¯ ∼ N (μ, σ 2 /n). bound and sometimes it is not necessarily attained. If the Xi ’s are normal, X 2 ¯ Hence, X is unbiased for μ with variance σ /n equal to the Cram´er-Rao lower bound. Therefore, ¯ is MVU for μ. On the other hand, X  ¯ 2 /n, σ 2M LE = ni=1 (Xi − X) and it can be shown that (n σ 2M LE )/(n − 1) = s2 is unbiased for σ 2 . In fact, (n − 1)s2 /σ 2 ∼ χ2n−1 and the expected value of a Chi-squared variable with (n − 1) degrees of freedom is exactly its degrees of freedom. Using this fact, E{(n − 1)s2 /σ 2 } = E(χ2n−1 ) = n − 1. Therefore, E(s2 ) = σ 2 .1 Also, the variance of a Chi-squared variable with (n − 1) degrees of freedom is twice these degrees of freedom. Using this fact, var{(n − 1)s2 /σ 2 } = var(χ2n−1 ) = 2(n − 1)


18

Chapter 2: Basic Statistical Concepts

or {(n − 1)2 /σ 4 }var(s2 ) = 2(n − 1). Hence, the var(s2 ) = 2σ 4 /(n − 1) and this does not attain the Cram´er-Rao lower bound. In fact, σ 2M LE ) = {(n − 1)2 /n2 }var(s2 ) = {2(n − 1)}σ 4 /n2 . it is larger than (2σ 4 /n). Note also that var( 4 This is smaller than (2σ /n)! How can that be? Remember that σ 2M LE is a biased estimator 2 2 of σ and hence, var( σ M LE ) should not be compared with the Cram´er-Rao lower bound. This lower bound pertains only to unbiased estimators. Warning: Attaining the Cram´er-Rao lower bound is only a sufficient condition for efficiency. Failing to satisfy this condition does not necessarily imply that the estimator is not efficient. Example 2: For the Bernoulli case logf (Xi ; θ) = Xi logθ + (1 − Xi )log(1 − θ) ∂logf (Xi , θ)/∂θ = (Xi /θ) − (1 − Xi )/(1 − θ) ∂ 2 logf (Xi ; θ)/∂θ 2 = (−Xi /θ 2 ) − (1 − Xi )/(1 − θ)2 and E[∂ 2 logf (Xi ; θ)/∂θ 2 ] = (−1/θ) − 1/(1 − θ) = −1/[θ(1 − θ)]. Therefore, for any unbiased estimator of θ, say  θ, its variance satisfies the following property: var( θ) ≥ θ(1 − θ)/n.

For the Bernoulli random sample, we proved that μ = E(Xi ) = θ. Similarly, it can be easily ¯ has mean μ = θ and var(X) ¯ = σ 2 /n = θ(1−θ)/n. verified that σ 2 = var(Xi ) = θ(1−θ). Hence, X ¯ is unbiased for θ and it attains the Cram´er-Rao lower bound. Therefore, X ¯ This means that X is MVU for θ. Unbiasedness and efficiency are finite sample properties (in other words, true for any finite sample size n). Once we let n tend to ∞ then we are in the realm of asymptotic properties.

Example 3: For a random sample from any distribution with mean μ it is clear that μ  = ¯ + 1/n) is not an unbiased estimator of μ since E( ¯ + 1/n) = μ + 1/n. However, as (X μ) = E(X n → ∞ the lim E( μ) is equal to μ. We say, that μ  is asymptotically unbiased for μ. Example 4: For the Normal case σ 2M LE = (n − 1)s2 /n

and E( σ 2M LE ) = (n − 1)σ 2 /n.

2M LE is asymptotically unbiased for σ 2 . But as n → ∞, lim E( σ 2M LE ) = σ 2 . Hence, σ Similarly, an estimator which attains the Cram´er-Rao lower bound in the limit is asymp¯ = σ 2 /n, and this tends to zero as n → ∞. Hence, we totically efficient. Note that var(X) √ ¯ √ ¯ ¯ = σ 2 . We say that the = n var(X) consider nX which has finite variance since var( nX) 2 ¯ ¯ asymptotic variance of X denoted by asymp.var(X) = σ /n and that it attains the Cram´er¯ is therefore asymptotically efficient. Similarly, Rao lower bound in the limit. X √ 2 var( n σ M LE ) = n var( σ 2M LE ) = 2(n − 1)σ 4 /n which tends to 2σ 4 as n → ∞. This means that asymp.var( σ 2M LE ) = 2σ 4 /n and that it attains 2 the Cram´er-Rao lower bound in the limit. Therefore, σ M LE is asymptotically efficient.


2.3

Properties of Estimators

19

(iii) Consistency ¯ − μ| > c] = 0 Another asymptotic property is consistency. This says that as n → ∞ lim Pr[|X ¯ for any arbitrary positive constant c. In other words, X will not differ from μ as n → ∞. Proving this property uses the Chebyshev’s inequality which states in this context that ¯ − μ| > kσ ¯ ] ≤ 1/k2 . Pr[|X X If we let c = kσ X¯ then 1/k2 = σ 2X¯ /c2 = σ 2 /nc2 and this tends to 0 as n → ∞, since σ 2 and c are finite positive constants. A sufficient condition for an estimator to be consistent is that it is asymptotically unbiased and that its variance tends to zero as n → ∞.2 ¯ =μ Example 1: For a random sample from any distribution with mean μ and variance σ 2 , E(X) ¯ = σ 2 /n → 0 as n → ∞, hence X ¯ is consistent for μ. and var(X)

Example 2: For the Normal case, we have shown that E(s2 ) = σ 2 and var(s2 ) = 2(n−1)σ 4 /n2 → 0 as n → ∞, hence s2 is consistent for σ 2 . ¯ = θ and var(X) ¯ = θ(1 − θ)/n → 0 as Example 3: For the Bernoulli case, we know that E(X) ¯ is consistent for θ. n → ∞, hence X Warning: This is only a sufficient condition for consistency. Failing to satisfy this condition does not necessarily imply that the estimator is inconsistent. (iv) Sufficiency ¯ is sufficient for μ, if X ¯ contains all the information in the sample pertaining to μ. In other X ¯ is independent of μ. To prove this fact one uses the factorization words, f (X1 , . . . , Xn /X) ¯ is sufficient for μ, if and only if one can theorem due to Fisher and Neyman. In this context, X factorize the joint p.d.f. ¯ μ) · g(X1 , . . . , Xn ) f (X1 , . . . , Xn ; μ) = h(X; where h and g are any two functions with the latter being only a function of the X’s and independent of μ in form and in the domain of the X’s. Example 1: For the Normal case, it is clear from equation (2.1) that by subtracting and adding ¯ in the summation we can write after some algebra X n 2 2 2 ¯ 2 ¯ f (X1 , .., Xn ; μ, σ 2 ) = (1/2πσ 2 )n/2 e−{(1/2σ ) i=1 (Xi −X) } e−{(n/2σ )(X−μ) } 2

¯

2

¯ μ) = e−(n/2σ )(X−μ ) and g(X1 , . . . , Xn ) is the remainder term which is independent Hence, h(X; ¯ is of μ in form. Also −∞ < Xi < ∞ and hence independent of μ in the domain. Therefore, X sufficient for μ. Example 2: For the Bernoulli case, ¯

¯

f (X1 , . . . , Xn ; θ) = θnX (1 − θ)n(1−X)

Xi = 0, 1

for i = 1, . . . , n.

¯ ¯ θ) = θ nX¯ (1 − θ)n(1−X) Therefore, h(X, and g(X1 , . . . , Xn ) = 1 which is independent of θ in form ¯ is sufficient for θ. and domain. Hence, X


20

Chapter 2: Basic Statistical Concepts

Under certain regularity conditions on the distributions we are sampling from, one can show that the MVU of any parameter θ is an unbiased function of a sufficient statistic for θ.3 Advantages of the maximum likelihood estimators is that (i) they are sufficient estimators when they exist. (ii) They are asymptotically efficient. (iii) If the distribution of the MLE satisfies certain regularity conditions, then making the MLE unbiased results in a unique MVU estimator. A prime example of this is s2 which was shown to be an unbiased estimator of σ 2 for a random sample drawn from the Normal distribution. It can be shown that s2 is sufficient for σ 2 and that (n − 1)s2 /σ 2 ∼ χ2n−1 . Hence, s2 is an unbiased sufficient statistic for σ 2 and therefore it is MVU for σ 2 , even though it does not attain the Cram´er-Rao lower bound. (iv) Maximum likelihood estimates are invariant with respect to continuous transformations. To explain the last property, ¯ an obvious estimator is eμ M LE = eX¯ . This is in consider the estimator of eμ . Given μ M LE = X, μ fact the MLE of e . In general, if g(μ) is a continuous function of μ, then g( μM LE ) is the MLE of g(μ). Note that E(eμ M LE ) = eE(μM LE ) = eμ , in other words, expectations are not invariant to all continuous transformations, especially nonlinear ones and hence the resulting MLE estimator ¯ ¯ is unbiased for μ. may not be unbiased. eX is not unbiased for eμ even though X In summary, there are two routes for finding the MVU estimator. One is systematically following the derivation of a sufficient statistic, proving that its distribution satisfies certain regularity conditions, and then making it unbiased for the parameter in question. Of course, MLE provides us with sufficient statistics, for example, ¯ M LE = X X1 , . . . , Xn ∼ IIN(μ, σ 2 ) ⇒ μ

and σ 2M LE =

n

i=1 (Xi

¯ 2 /n − X)

¯ is unbiased for μ and X ¯ ∼ N (μ, σ 2 /n). The are both sufficient for μ and σ 2 , respectively. X ¯ to be MVU for μ. σ Normal distribution satisfies the regularity conditions needed for X 2M LE is 2 2 2 2 2 2 2 biased for σ , but s = n σ M LE /(n − 1) is unbiased for σ and (n − 1)s /σ ∼ χn−1 which also satisfies the regularity conditions for s2 to be a MVU estimator for σ 2 . Alternatively, one finds the Cram´er-Rao lower bound and checks whether the usual estimator (obtained from say the method of moments or the maximum likelihood method) achieves this lower bound. If it does, this estimator is efficient, and there is no need to search further. If it does not, the former strategy leads us to the MVU estimator. In fact, in the previous example ¯ attains the Cram´er-Rao lower bound, whereas s2 does not. However, both are MVU for μ X and σ 2 respectively.

(v) Comparing Biased and Unbiased Estimators Suppose we are given two estimators  θ1 and  θ2 of θ where the first is unbiased and has a large variance and the second is biased but with a small variance. The question is which one of these two estimators is preferable?  θ1 is unbiased whereas  θ2 is biased. This means that if we repeat the sampling procedure many times then we expect  θ1 to be on the average correct, whereas  θ2 would be on the average different from θ. However, in real life, we observe only one sample. With a large variance for  θ 1 , there is a great likelihood that the sample drawn could result in a θ1 far away from θ. However, with a small variance for  θ2 , there is a better chance of getting a θ2 close to θ. If our loss function is L( θ, θ) = ( θ − θ)2 then our risk is R( θ, θ) = E[L( θ, θ)] = E( θ − θ)2 = M SE( θ) 2     = E[θ − E(θ) + E(θ) − θ] = var(θ) + (Bias( θ))2 .


2.4

Hypothesis Testing

21

Minimizing the risk when the loss function is quadratic is equivalent to minimizing the Mean Square Error (MSE). From its definition the MSE shows the trade-off between bias and variance. MVU theory, sets the bias equal to zero and minimizes var( θ). In other words, it minimizes the above risk function but only over  θ’s that are unbiased. If we do not restrict ourselves to unbiased estimators of θ, minimizing MSE may result in a biased estimator such as  θ2 which beats  θ1 because the gain from its smaller variance outweighs the loss from its small bias, see Figure 2.2. f (Gˆ )

f (Gˆ 2 )

Bias

f (Gˆ1 )

Gˆ G E (Gˆ 2 ) Figure 2.2 Bias versus Variance

2.4

Hypothesis Testing

The best way to proceed is with an example. Example 1: The Economics Departments instituted a new program to teach micro-principles. We would like to test the null hypothesis that 80% of economics undergraduate students will pass the micro-principles course versus the alternative hypothesis that only 50% will pass. We draw a random sample of size 20 from the large undergraduate micro-principles class and as a simple rule we accept the null if x, the number of passing students is larger or equal to 13, otherwise the alternative hypothesis will be accepted. Note that the distribution we are drawing from is Bernoulli with the probability of success θ, and we have chosen only two states of the world H0 ; θ0 = 0.80 and H1 ; θ 1 = 0.5. This situation is known as testing a simple hypothesis versus another simple hypothesis because the distribution is completely specified under the null or alternative hypothesis. One would expect (E(x) = nθ0 ) 16 students under H0 and (nθ1 ) 10 students under H1 to pass the micro-principles exams. It seems then logical to take x ≥ 13 as the cut-off point distinguishing H0 form H1 . No theoretical justification is given at this stage to this arbitrary choice except to say that it is the mid-point of [10, 16]. Figure 2.3 shows that one can make two types of errors. The first is rejecting H0 when in fact it is true, this is known as type I error and the probability of committing this error is denoted by α. The second is accepting H1 when it is false. This is known as type II error and the corresponding probability is denoted by β. For this example


22

Chapter 2: Basic Statistical Concepts

α = Pr[rejecting H0 /H0 is true] = Pr[x < 13/θ = 0.8] = b(n = 20; x = 0; θ = 0.8) + .. + b(n = 20; x = 12; θ = 0.8) = b(n = 20; x = 20; θ = 0.2) + .. + b(n = 20; x = 8; θ = 0.2) = 0 + .. + 0 + 0.0001 + 0.0005 + 0.0020 + 0.0074 + 0.0222 = 0.0322 where we have used the fact that b(n; x; θ) = b(n; n − x; 1 − θ) and b(n; x; θ) = denotes the binomial distribution for x = 0, 1, . . . , n, see problem 4.

 n x n−x x θ (1 − θ)

True World

Decision

θ 0 = 0.80

θ1 = 0.50

θ0

No error

Type II error

θ1

Type I error

No Error

Figure 2.3 Type I and II Error

β = Pr[accepting H0 /H0 is false] = Pr[x ≥ 13/θ = 0.5]

= b(n = 20; x = 13; θ = 0.5) + .. + b(n = 20; x = 20; θ = 0.5) = 0.0739 + 0.0370 + 0.0148 + 0.0046 + 0.0011 + 0.0002 + 0 + 0 = 0.1316

The rejection region for H0 , x < 13, is known as the critical region of the test and α = Pr[Falling in the critical region/H0 is true] is also known as the size of the critical region. A good test is one which minimizes both types of errors α and β. For the above example, α is low but β is high with more than a 13% chance of happening. This β can be reduced by changing the critical region from x < 13 to x < 14, so that H0 is accepted only if x ≥ 14. In this case, one can easily verify that α = Pr[x < 14/θ = 0.8] = b(n = 20; x = 0; θ = 0.8) + .. + b(n = 20, x = 13, θ = 0.8) = 0.0322 + b(n = 20; x = 13; θ = 0.8) = 0.0322 + 0.0545 = 0.0867 and β = Pr[x ≥ 14/θ = 0.5] = b(n = 20; x = 14; θ = 0.5) + .. + b(n = 20; x = 20; θ = 0.5) = 0.1316 − b(n = 20; x = 13; θ = 0.5) = 0.0577

By becoming more conservative on accepting H0 and more liberal on accepting H1 , one reduces β from 0.1316 to 0.0577 but the price paid is the increase in α from 0.0322 to 0.0867. The only way to reduce both α and β is by increasing n. For a fixed n, there is a tradeoff between α and β as we change the critical region. To understand this clearly, consider the real life situation of trial by jury for which the defendant can be innocent or guilty. The decision of incarceration or release implies two types of errors. One can make α = Pr[incarcerating/innocence] = 0 and β = its maximum, by releasing every defendant. Or one can make β = Pr[release/guilty] = 0 and α = its maximum, by incarcerating every defendant. These are extreme cases but hopefully they demonstrate the trade-off between α and β.


2.4

Hypothesis Testing

23

The Neyman-Pearson Theory The classical theory of hypothesis testing, known as the Neyman-Pearson theory, fixes α = Pr(type I error) ≤ a constant and minimizes β or maximizes (1 − β). The latter is known as the Power of the test under the alternative. The Neyman-Pearson Lemma: If C is a critical region of size α and k is a constant such that (L0 /L1 ) ≤ k inside C and (L0 /L1 ) ≥ k outside C then C is a most powerful critical region of size α for testing H0 ; θ = θ0 , against H1 ; θ = θ1 . Note that the likelihood has to be completely specified under the null and alternative. Hence, this lemma applies only to testing a simple versus another simple hypothesis. The proof of this lemma is given in Freund (1992). Intuitively, L0 is the likelihood function under the null H0 and L1 is the corresponding likelihood function under H1 . Therefore, (L0 /L1 ) should be small for points inside the critical region C and large for points outside the critical region C. The proof of the theorem shows that any other critical region, say D, of size α cannot have a smaller probability of type II error than C. Therefore, C is the best or most powerful critical region of size α. Its power (1 − β) is maximum at H1 . Let us demonstrate this lemma with an example. Example 2: Given a random sample of size n from N (μ, σ 2 = 4), use the Neyman-Pearson lemma to find the most powerful critical region of size α = 0.05 for testing H0 ; μ0 = 2 against the alternative H1 ; μ1 = 4. Note that this is a simple versus simple hypothesis as required by the lemma, since σ 2 = 4 is known and μ is specified by H0 and H1 . The likelihood function for the N (μ, 4) density is given by √    L(μ) = f (x1 , . . . , xn ; μ, 4) = (1/2 2π)n exp − ni=1 (xi − μ)2 /8

so that

and

√    L0 = L(μ0 ) = (1/2 2π)n exp − ni=1 (xi − 2)2 /8 √    L1 = L(μ1 ) = (1/2 2π)n exp − ni=1 (xi − 4)2 /8

Therefore

 n  n n 2 2 L0 /L1 = exp − i=1 (xi − 2) − i=1 (xi − 4) /8 = exp {− i=1 xi /2 + 3n/2}

and the critical region is defined by  exp {− ni=1 xi /2 + 3n/2} ≤ k

inside C

Taking logarithms of both sides, subtracting (3/2)n and dividing by (−1/2)n one gets x ¯≥K

inside C


24

Chapter 2: Basic Statistical Concepts

In practice, one need not keep track of K as long as one keeps track of the direction of the inequality. K can be determined by making the size of C = α = 0.05. In this case √ α = Pr[¯ x ≥ K/μ = 2] = Pr[z ≥ (K − 2)/(2/ n)]

√ where z = (¯ x − 2)/(2/ n) is distributed N (0, 1) under H0 . From the N (0, 1) tables, we have K −2 √ = 1.645 (2/ n) Hence, √ K = 2 + 1.645(2/ n) √ and x ¯ ≥ 2 + 1.645(2/ n) defines the most powerful critical region of size α = 0.05 for testing H0 ; μ0 = 2 versus H1 ; μ1 = 4. Note that, in this case √ β = Pr[¯ x < 2 + 1.645(2/ n)/μ = 4] √ √ √ = Pr[z < [−2 + 1.645(2/ n)]/(2/ n)] = Pr[z < 1.645 − n] For n = 4; β = Pr[z < −0.355] = 0.3613 shown by the shaded region in Figure 2.4. For n = 9; β = Pr[z < −1.355] = 0.0877, and for n = 16; β = Pr[z < −2.355] = 0.00925.

a = 0.05

b = 0.3613 m0 = 2

3.645

m1 = 4

x

Figure 2.4 Critical Region for Testing μ0 = 2 against μ1 = 4 for n = 4

This gives us an idea of how, for a fixed α = 0.05, the minimum β decreases with larger sample size n. As n increases from 4 to 9 to 16, the var(¯ x) = σ 2 /n decreases and the two distributions shown in Figure 2.4 shrink in dispersion still centered around μ0 = 2 and μ1 = 4, respectively. This allows better decision making (based on larger sample size) as reflected by the critical region shrinking from x ≥ 3.65 for n = 4 to x ≥ 2.8225 for n = 16, and the power (1 − β) rising from 0.6387 to 0.9908, respectively, for a fixed α ≤ 0.05. The power function is the probability of rejecting H0 . It is equal to α under H0 and 1 − β under H1 . The ideal power function is zero at H0 and one at H1 . The Neyman-Pearson lemma allows us to fix α, say at 0.05, and find the test with the best power at H1 . In example 2, both the null and alternative hypotheses are simple. In real life, one is more likely to be faced with testing H0 ; μ = 2 versus H1 ; μ = 2. Under the alternative hypothesis, the distribution is not completely specified, since the mean μ is not known, and this is referred to as a composite hypothesis. In this case, one cannot compute the probability of type II error


2.4

Hypothesis Testing

25

since the distribution is not known under the alternative. Also, the Neyman-Pearson lemma cannot be applied. However, a simple generalization allows us to compute a Likelihood Ratio test which has satisfactory properties but is no longer uniformly most powerful of size α. In this case, one replaces L1 , which is not known since H1 is a composite hypothesis, by the maximum value of the likelihood, i.e., λ=

maxL0 maxL

Since max L0 is the maximum value of the likelihood under the null while maxL is the maximum value of the likelihood over the whole parameter space, it follows that maxL0 ≤ maxL and λ ≤ 1. Hence, if H0 is true, λ is close to 1, otherwise it is smaller than 1. Therefore, λ ≤ k defines the critical region for the Likelihood Ratio test, and k is determined such that the size of this test is α. Example 3: For a random sample x1 , . . . , xn drawn from a Normal distribution with mean μ and variance σ 2 = 4, derive the Likelihood Ratio test for H0 ; μ = 2 versus H1 ; μ = 2. In this case √    maxL0 = (1/2 2π)n exp − ni=1 (xi − 2)2 /8 = L0 and

√    ¯)2 /8 = L( μM LE ) maxL = (1/2 2π)n exp − ni=1 (xi − x

where use is made of the fact that μ M LE = x ¯. Therefore,       λ = exp − ni=1 (xi − 2)2 + ni=1 (xi − x ¯)2 /8 = exp −n(¯ x − 2)2 /8

Hence, the region for which λ ≤ k, is equivalent after some simple algebra to the following region (¯ x − 2)2 ≥ K

or

|¯ x − 2| ≥ K 1/2

where K is determined such that Pr[|¯ x − 2| ≥ K 1/2 /μ = 2] = α

√ We know that x ¯ ∼ N (2, 4/n) under H0 . Hence, z = (¯ x − 2)/(2/ n) is N (0, 1) under H0 , and the critical region of size α will be based upon |z| ≥ zα/2 where zα/2 is given in Figure 2.5 and is the value of a N (0, 1) random variable such that the probability of exceeding it is α/2. For α = 0.05, zα/2 = 1.96, and for α = 0.10, zα/2 = 1.645. This is a two-tailed test with rejection of H0 obtained in case z ≤ −zα/2 or z ≥ zα/2 . Note that in this case LR ≡ −2logλ = (¯ x − 2)2 /(4/n) = z 2 which is distributed as χ21 under H0 . This is because it is the square of a N (0, 1) random variable under H0 . This is a finite sample result holding for any n. In general, other examples may lead to more complicated λ statistics for which it is difficult to find the corresponding distributions and hence the corresponding critical values. For these cases, we have an asymptotic result


26

Chapter 2: Basic Statistical Concepts

Reject Ho

Reject Ho

Do not reject Ho

=/2

=/2

– z= / 2

z= / 2

0

Figure 2.5 Critical Values

which states that, for large n, LR = −2logλ will be asymptotically distributed as χ2ν where ν denotes the number of restrictions that are tested by H0 . For example 2, ν = 1 and hence, LR is asymptotically distributed as χ21 . Note that we did not need this result as we found LR is exactly distributed as χ21 for any n. If one is testing H0 ; μ = 2 and σ 2 = 4 against the alternative that H1 ; μ = 2 or σ 2 = 4, then the corresponding LR will be asymptotically distributed as χ22 , see problem 5, part (f). Likelihood Ratio, Wald and Lagrange Multiplier Tests Before we go into the derivations of these three tests we start by giving an intuitive graphical explanation that will hopefully emphasize the differences among these tests. This intuitive explanation is based on the article by Buse (1982). Consider a quadratic log-likelihood function in a parameter of interest, say μ. Figure 2.6 shows this log-likelihood logL(μ), with a maximum at μ . The Likelihood Ratio test, tests the μ) where null hypothesis H0 ; μ = μ0 by looking at the ratio of the likelihoods λ = L(μ0 )/L( log L( m )

log L( mˆ )

A.

log L1 ( m 0 )

B.

log L2 ( m 0 )

C.

log L1 ( m )

log L2 ( m ) m0 Figure 2.6 Wald Test

m


2.4

Hypothesis Testing

27

−2logλ, twice the difference in log-likelihood, is distributed asymptotically as χ21 under H0 . This test differentiates between the top of the hill and a preassigned point on the hill by evaluating the height at both points. Therefore, it needs both the restricted and unrestricted maximum of the likelihood. This ratio is dependent on the distance of μ0 from μ  and the curvature of . In fact, for a fixed ( μ − μ0 ), the larger C( μ), the log-likelihood, C(μ) = |∂ 2 logL(μ)/∂μ2 |, at μ the larger is the difference between the two heights. Also, for a given curvature at μ , the larger ( μ − μ0 ) the larger is the difference between the heights. The Wald test works from the top of the hill, i.e., it needs only the unrestricted maximum likelihood. It tries to establish the distance μ − μ0 ), and the curvature at μ . In fact the Wald to μ0 , by looking at the horizontal distance ( statistic is W = ( μ − μ0 )2 C( μ) and this is asymptotically distributed as χ21 under H0 . The usual form of W has I(μ) = −E[∂ 2 logL(μ)/∂μ2 ] the information matrix evaluated at μ , rather than C( μ), but the latter is a consistent estimator of I(μ). The information matrix will be studied in details in Chapter 7. It will be shown, under fairly general conditions, that μ  the MLE of μ − μ0 )2 /var( μ) all evaluated at the unrestricted MLE. μ, has var( μ) = I −1 (μ). Hence W = ( The Lagrange-Multiplier test (LM), on the other hand, goes to the preassigned point μ0 , i.e., it only needs the restricted maximum likelihood, and tries to determine how far it is from the top of the hill by considering the slope of the tangent to the likelihood S(μ) = ∂logL(μ)/∂μ at μ0 , and the rate at which this slope is changing, i.e., the curvature at μ0 . As Figure 2.7 shows, for two log-likelihoods with the same S(μ0 ), the one that is closer to the top of the hill is the one with the larger curvature at μ0 . log L( m )



log L1 ( mˆ 1 )

log L1 ( m )

log L2 ( mˆ 2 ) log L( m0 ) log L2 ( m )

m m0

mˆ 2

mˆ 1

Figure 2.7 LM Test

This suggests the following statistic: LM = S 2 (μ0 ){C(μ0 )}−1 where the curvature appears in inverse form. In the Appendix to this chapter, we show that the E[S(μ)] = 0 and var[S(μ)] = I(μ). Hence LM = S 2 (μ0 )I −1 (μ0 ) = S 2 (μ0 )/var[S(μ0 )] all evaluated at the restricted MLE. Another interpretation of the LM test is that it is a measure of failure of the restricted estimator, in this case μ0 , to satisfy the first-order conditions of maximization of the unrestricted likelihood. We know that S( μ) = 0. The question is: to what extent does S(μ0 ) differ from zero? S(μ) is known in the statistics literature as the score, and the LM test is also referred to as the score test. For a more formal treatment of these tests, let us reconsider example 3 of a random sample x1 , . . . , xn from a N (μ, 4) where we are interested in testing H0 ; μ0 = 2 versus H1 ; μ = 2. The likelihood function L(μ) as well as LR = −2logλ = n(¯ x − 2)2 /4 were given in example 3. In


28

Chapter 2: Basic Statistical Concepts

fact, the score function is given by n (xi â&#x2C6;&#x2019; Îź) n(ÂŻ x â&#x2C6;&#x2019; Îź) â&#x2C6;&#x201A;logL(Îź) = i=1 = S(Îź) = â&#x2C6;&#x201A;Îź 4 4 and under H0 n(ÂŻ x â&#x2C6;&#x2019; 2) 4 â&#x2C6;&#x201A; 2 logL(Îź) n n C(Îź) = | | =|â&#x2C6;&#x2019; |= â&#x2C6;&#x201A;Îź2 4 4

2 n â&#x2C6;&#x201A; logL(Îź) = = C(Îź). and I(Îź) = â&#x2C6;&#x2019;E 2 â&#x2C6;&#x201A;Îź 4 The Wald statistic is based on S(Îź0 ) = S(2) =

W = ( ÎźM LE â&#x2C6;&#x2019; 2)2 I( ÎźM LE ) = (ÂŻ x â&#x2C6;&#x2019; 2)2 ¡ The LM statistic is based on LM = S 2 (Îź0 )I â&#x2C6;&#x2019;1 (Îź0 ) =

n 4

n2 (ÂŻ x â&#x2C6;&#x2019; 2)2 4 n(ÂŻ x â&#x2C6;&#x2019; 2)2 ¡ = 16 n 4

Therefore, W = LM = LR for this example with known variance Ď&#x192; 2 = 4. These tests are all based upon the |ÂŻ x â&#x2C6;&#x2019; 2| â&#x2030;Ľ k critical region, where k is determined such that the size of the test is Îą. In general, these test statistics are not always equal, as is shown in the next example. Example 4: For a random sample x1 , . . . , xn drawn from a N (Îź, Ď&#x192; 2 ) with unknown Ď&#x192; 2 , test the hypothesis H0 ; Îź = 2 versus H1 ; Îź = 2. Problem 5, part (c), asks the reader to verify that

n 2 n2 (ÂŻ x â&#x2C6;&#x2019; 2)2 n2 (ÂŻ x â&#x2C6;&#x2019; 2)2 i=1 (xi â&#x2C6;&#x2019; 2)    whereas W = and LM = . LR = nlog n n n 2 ÂŻ)2 ÂŻ)2 i=1 (xi â&#x2C6;&#x2019; x i=1 (xi â&#x2C6;&#x2019; x i=1 (xi â&#x2C6;&#x2019; 2)

One can easily show that LM/n = (W/n)/[1+(W/n)] and LR/n = log[1+(W/n)]. Let y = W/n, then using the inequality y â&#x2030;Ľ log(1 + y) â&#x2030;Ľ y/(1 + y), one can conclude that W â&#x2030;Ľ LR â&#x2030;Ľ LM . This inequality was derived by Berndt and Savin (1977), and will be considered again when we study test of hypotheses in the general linear model. Note, however that all three test statistics are based upon |ÂŻ x â&#x2C6;&#x2019; 2| â&#x2030;Ľ k and for ďŹ nite n, the same exact critical value could be obtained from the Normally distributed x ÂŻ. This section introduced the W, LR and LM test statistics, all of which have the same asymptotic distribution. In addition, we showed that using the normal distribution, when Ď&#x192; 2 is known, W = LR = LM for testing H0 ; Îź = 2 versus H1 ; Îź = 2. However, when Ď&#x192; 2 is unknown, we showed that W â&#x2030;Ľ LR â&#x2030;Ľ LM for the same hypothesis. Example 5: For a random sample x1 , . . . , xn drawn from a Bernoulli distribution with parameter θ, test the hypothesis H0 ; θ = θ 0 versus H1 ; θ = θ0 , where θ0 is a known positive fraction. This example is based on Engle (1984). Problem 4, part (i), asks the reader to derive LR, W and LM for H0 ; θ = 0.2 versus H1 ; θ = 0.2. The likelihood L(θ) and the Score S(θ) were derived in section 2.2. One can easily verify that n  n â&#x2C6;&#x2019; ni=1 xi â&#x2C6;&#x201A; 2 logL(θ) i=1 xi C(θ) = | |= + (1 â&#x2C6;&#x2019; θ)2 â&#x2C6;&#x201A;θ 2 θ2


2.4

and

Hypothesis Testing

29

n ∂ 2 logL(θ) = I(θ) = −E 2 θ(1 − θ) ∂θ

The Wald statistic is based on W = ( θ M LE − θ0 )2 I( θ M LE ) = (¯ x − θ0 )2 ·

n (¯ x − θ0 )2 = x ¯(1 − x ¯) x ¯(1 − x ¯)/n

using the fact that  θM LE = x ¯. The LM statistic is based on LM = S 2 (θ0 )I −1 (θ0 ) =

θ0 (1 − θ0 ) (¯ x − θ0 )2 (¯ x − θ0 )2 · = [θ 0 (1 − θ0 )/n]2 n θ0 (1 − θ 0 )/n

Note that the numerator of the W and LM are the same. It is the denominator which is the var(¯ x) = θ(1 − θ)/n that is different. For Wald, this var(¯ x) is evaluated at  θM LE , whereas for LM, this is evaluated at θ0 . The LR statistic is based on   logL( θ M LE ) = ni=1 xi log¯ x + (n − ni=1 xi )log(1 − x ¯) and

logL(θ 0 ) = so that

n

i=1 xi logθ 0

+ (n −

n

i=1 xi )log(1

− θ0)

 θM LE ) = −2[ ni=1 xi (logθ0 − log¯ x) LR = −2logL(θ0 ) + 2logL( n +(n − i=1 xi )(log(1 − θ0 ) − log(1 − x ¯))]

For this example, LR looks different from W and LM. However, a second-order Taylor-Series expansion of LR around θ0 = x ¯ yields the same statistic. Also, for n → ∞, plim x ¯ = θ and if H0 is true, then all three statistics are asymptotically equivalent. Note also that all three test statistics are based upon |¯ x − θ0 | ≥ k and for finite n, the same exact critical value could be obtained from the binomial distribution. See problem 19 for more examples of the conflict in test of hypotheses using the W, LR and LM test statistics. Bera and Permaratne (2001, p. 58) tell the following amusing story that can bring home the interrelationship among the three tests: “Once around 1946 Ronald Fisher invited Jerzy Neyman, Abraham Wald, and C.R. Rao to his lodge for afternoon tea. During their conversation, Fisher mentioned the problem of deciding whether his dog, who had been going to an “obedience school” for some time, was disciplined enough. Neyman quickly came up with an idea: leave the dog free for some time and then put him on his leash. If there is not much difference in his behavior, the dog can be thought of as having completed the course successfully. Wald, who lost his family in the concentration camps, was adverse to any restrictions and simply suggested leaving the dog free and seeing whether it behaved properly. Rao, who had observed the nuisances of stray dogs in Calcutta streets did not like the idea of letting the dog roam freely and suggested keeping the dog on a leash at all times and observing how hard it pulls on the leash. If it pulled too much, it needed more training. That night when Rao was back in his Cambridge dormitory after tending Fisher’s mice at the genetics laboratory, he suddenly realized the connection of Neyman and Wald’s recommendations to the Neyman-Pearson LR and Wald tests. He got an idea and the rest is history.”


30

Chapter 2: Basic Statistical Concepts

2.5

Confidence Intervals

Estimation methods considered in section 2.2 give us a point estimate of a parameter, say μ, and that is the best bet, given the data and the estimation method, of what μ might be. But it is always good policy to give the client an interval, rather than a point estimate, where with some degree of confidence, usually 95% confidence, we expect μ to lie. We have seen in Figure 2.5 that for a N (0, 1) random variable z, we have Pr[−zα/2 ≤ z ≤ zα/2 ] = 1 − α and for α = 5%, this probability is 0.95, giving the required 95% confidence. In fact, zα/2 = 1.96 and Pr[−1.96 ≤ z ≤ 1.96] = 0.95 This says that if we draw 100 random numbers from a N (0, 1) density, (using a normal random number generator) we expect 95 out of these 100 numbers to lie in the [−1.96, 1.96] interval. Now, let us get back to the problem of estimating μ from a random sample x1 , . . . , xn drawn M LE = x ¯ and x ¯ ∼ N (μ, σ 2 /n). Hence, from a N (μ, σ 2 ) distribution. We found out that μ √ ¯ observed from the sample, and the z = (¯ x − μ)/(σ/ n) is N (0, 1). The point estimate for μ is x 95% confidence interval for μ is obtained by replacing z by its value in the above probability statement: Pr[−zα/2 ≤

x ¯−μ √ ≤ zα/2 ] = 1 − α σ/ n

Assuming σ is known for the moment, one can rewrite this probability statement after some simple algebraic manipulations as √ √ ¯ + zα/2 (σ/ n)] = 1 − α Pr[¯ x − zα/2 (σ/ n) ≤ μ ≤ x Note that this probability statement has random variables on both ends and the probability that these random variables sandwich the unknown parameter μ is 1 − α. With the same confidence of drawing 100 random N (0, 1) numbers and finding 95 of them falling in the (−1.96, 1.96) range we are confident that if we drew a 100 samples and computed a 100 x ¯’s, and a 100 intervals √ (¯ x ± 1.96 σ/ n), μ will lie in these intervals in 95 out of 100 times. If σ is not known, and is replaced by s, then problem 12 shows that this is equivalent to dividing a N (0, 1) random variable by an independent χ2n−1 random variable divided by its degrees of freedom, leading to a t-distribution with (n − 1) degrees of freedom. Hence, using the t-tables for (n − 1) degrees of freedom Pr[−tα/2;n−1 ≤ tn−1 ≤ tα/2;n−1 ] = 1 − α √ and replacing tn−1 by (¯ x − μ)/(s/ n) one gets √ √ ¯ + tα/2;n−1 (s/ n)] = 1 − α Pr[¯ x − tα/2;n−1 (s/ n) ≤ μ ≤ x Note that the degrees of freedom (n−1) for the t-distribution come from s and the corresponding critical value tn−1;α/2 is therefore sample specific, unlike the corresponding case for the normal density where zα/2 does not depend on n. For small n, the tα/2 values differ drastically from


2.6

Descriptive Statistics

31

Table 2.1 Descriptive Statistics for the Earnings Data Sample: 1 595

Mean Median Maximum Minimum Std. Dev. Skewness Kurtosis Jarque-Bera Probability

Observations

LWAGE

WKS

ED

EX

MS

FEM

BLK

UNION

6.9507 6.9847 8.5370 5.6768 0.4384 –0.1140 3.3937

46.4520 48.0000 52.0000 5.0000 5.1850 –2.7309 13.7780

12.8450 12.0000 17.0000 4.0000 2.7900 –0.2581 2.7127

22.8540 21.0000 51.0000 7.0000 10.7900 0.4208 2.0086

0.8050 1.0000 1.0000 0.0000 0.3965 –1.5400 3.3715

0.1126 0.0000 1.0000 0.0000 0.3164 2.4510 7.0075

0.0723 0.0000 1.0000 0.0000 0.2592 3.3038 11.9150

0.3664 0.0000 1.0000 0.0000 0.4822 0.5546 1.3076

5.13 0.0769

3619.40 0.0000

8.65 0.0132

41.93 0.0000

238.59 0.0000

993.90 0.0000

3052.80 0.0000

101.51 0.0000

595

595

595

595

595

595

595

595

zα/2 , emphasizing the importance of using the t-density in small samples. When n is large the difference between zα/2 and tα/2 diminishes as the t-density becomes more like a normal density. For n = 20, and α = 0.05, tα/2;n−1 = 2.093 as compared with zα/2 = 1.96. Therefore, Pr[−2.093 ≤ tn−1 ≤ 2.093] = 0.95

√ and μ lies in x ¯ ± 2.093(s/ n) with 95% confidence. More examples of confidence intervals can be constructed, but the idea should be clear. Note that these confidence intervals are the other side of the coin for tests of hypotheses. For example, in testing H0 ; μ = 2 versus H1 ; μ = 2 for a known σ, we discovered that the Likelihood Ratio test is based on the same probability statement that generated the confidence interval for μ. In classical tests of hypothesis, we choose the level of confidence α = 5% and compute √ z = (¯ x − μ)/(σ/ n). This can be done since σ is known and μ = 2 under the null hypothesis H0 . Next, we do not reject H0 if z lies in the (−zα/2 , zα/2 ) interval and reject H0 otherwise. For confidence intervals, on the other hand, we do not know μ, and armed with a level of confidence (1 − α)% we construct the interval that should contain μ with that level of confidence. Having done that, if μ = 2 lies in that 95% confidence interval, then we cannot reject H0 ; μ = 2 at the 5% level. Otherwise, we reject H0 . This highlights the fact that any value of μ that lies in this 95% confidence interval (assuming it was our null hypothesis) cannot be rejected at the 5% level by this sample. This is why we do not say “accept H0 ”, but rather we say “do not reject H0 ”.

2.6

Descriptive Statistics

In Chapter 4, we will consider the estimation of a simple wage equation based on 595 individuals drawn from the Panel Study of Income Dynamics for 1982. This data is available on the Springer web site as EARN.ASC. Table 2.1 gives the descriptive statistics using EViews for a subset of the variables in this data set.


32

Chapter 2: Basic Statistical Concepts

80 Series: LWAGE Sample 1 595 Observations 595

70 60

Mean Median Maximum Minimum Std. Dev. Skewness Kurtosis

50 40 30 20 10

6.950745 6.984720 8.537000 5.676750 0.438403 â&#x20AC;&#x201C;0.114001 3.393651

Jarque-Bera Probability

0 6.0

6.5

7.0

7.5

8.0

5.130525 0.076899

8.5

Figure 2.8 Log (Wage) Histogram 250 Series: WKS Sample 1595 Observations 595

200

Mean Median Maximum Minimum Std. Dev. Skewness Kurtosis

150

100

46.45210 48.00000 52.00000 5.000000 5.185025 â&#x20AC;&#x201C;2.730880 13.77787

50 Jarque-Bera Probability

0 10

20

30

40

3619.416 0.000000

50

Figure 2.9 Weeks Worked Histogram

The average log wage is $6.95 for this sample with a minimum of $5.68 and a maximum of $8.54. The standard deviation of log wage is 0.44. A plot of the log wage histogram is given in Figure 2.8. Weeks worked vary between 5 and 52 with an average of 46.5 and a standard deviation of 5.2. This variable is highly skewed as evidenced by the histogram in Figure 2.9. Years of education vary between 4 and 17 with an average of 12.8 and a standard deviation of 2.79. There is the usual bunching up at 12 years, which is also the median, as is clear from Figure 2.10. Experience varies between 7 and 51 with an average of 22.9 and a standard deviation of 10.79. The distribution of this variable is skewed to the left, as shown in Figure 2.11. Marital status is a qualitative variable indicating whether the individual is married or not. This information is recoded as a numeric (1, 0) variable, one if the individual is married and zero


2.6

Descriptive Statistics

33

250 Series: ED Sample 1595 Observation 595

200

Mean Median Maximum Minimum Std. Dev. Skewness Kurtosis

150

100

12.84538 12.00000 17.00000 4.000000 2.790006 â&#x20AC;&#x201C;0.258116 2.712730

50 Jarque-Bera Probability

0 4

6

8

10

12

14

8.652780 0.013215

16

Figure 2.10 Years of Education Histogram 80 Series: EX Sample 1 595 Observation 595

70 60 50 40 30 20 10 0 10

20

30

40

Mean Median Maximum Minimum Std. Dev. Skewness Kurtosis

22.85378 21.00000 51.00000 7.000000 10.79018 0.420826 2.008578

Jarque-Bera Probability

41.93007 0.000000

50

Figure 2.11 Years of Experience Histogram

otherwise. This recoded variable is also known as a dummy variable. It is basically a switch turning on when the individual is married and off when he or she is not. Female is another dummy variable taking the value one when the individual is a female and zero otherwise. Black is a dummy variable taking the value one when the individual is black and zero otherwise. Union is a dummy variable taking the value one if the individualâ&#x20AC;&#x2122;s wage is set by a union contract and zero otherwise. The minimum and maximum values for these dummy variables are obvious. But if they were not zero and one, respectively, you know that something is wrong. The average is a meaningful statistic indicating the percentage of married individuals, females, blacks and union contracted wages in the sample. These are 80.5, 11.3, 7.2 and 30.6%, respectively. We would like to investigate the following claims: (i) women are paid less than men; (ii) blacks are paid less than non-blacks; (iii) married individuals earn more than non-married individuals; and (iv) union contracted wages are higher than non-union wages.


34

Chapter 2: Basic Statistical Concepts

Table 2.2 Test for the Difference in Means Average log wage Male Female Non-Black Black Not Married Married Non-Union Union

Difference

$7,004 $6,530 $6,978 $6,601 $6,664 $7,020 $6,945 $6,962

–0.474 (–8.86) –0.377 (–5.57) 0.356 (8.28) 0.017 (0.45)

Table 2.3 Correlation Matrix

LWAGE WKS ED EX MS FEM BLK UNION

LWAGE

WKS

ED

EX

MS

FEM

BLK

UNION

1.0000 0.0403 0.4566 0.0873 0.3218 –0.3419 –0.2229 0.0183

0.0403 1.0000 0.0002 –0.1061 0.0782 –0.0875 –0.0594 –0.1721

0.4566 0.0002 1.0000 –0.2219 0.0184 –0.0012 –0.1196 –0.2719

0.0873 –0.1061 –0.2219 1.0000 0.1570 –0.0938 0.0411 0.0689

0.3218 0.0782 0.0184 0.1570 1.0000 –0.7104 –0.2231 0.1189

–0.3419 –0.0875 –0.0012 –0.0938 –0.7104 1.0000 0.2086 –0.1274

–0.2229 –0.0594 –0.1196 0.0411 –0.2231 0.2086 1.0000 0.0302

0.0183 –0.1721 –0.2719 0.0689 0.1189 –0.1274 0.0302 1.0000

A simple first check could be based on computing the average log wage for each of these categories and testing whether the difference in means is significantly different from zero. This can be done using a t-test, see Table 2.2. The average log wage for males and females is given along with their difference and the corresponding t-statistic for the significance of this difference. Other rows of Table 2.2 give similar statistics for other groupings. In Chapter 4, we will show that this t-test can be obtained from a simple regression of log wage on the categorical dummy variable distinguishing the two groups. In this case, the Female dummy variable. From Table 2.2, it is clear that only the difference between union and non-union contracted wages are insignificant. One can also plot log wage versus experience, see Figure 2.12, log wage versus education, see Figure 2.13, and log wage versus weeks, see Figure 2.14. The data shows that, in general, log wage increases with education level, weeks worked, but that it exhibits a rising and then a declining pattern with more years of experience. Note that the t-tests based on the difference in log wage across two groupings of individuals, by sex, race or marital status, or the figures plotting log wage versus education, log wage versus weeks worked are based on pairs of variables in each case. A nice summary statistic based also on pairwise comparisons of these variables is the correlation matrix across the data. This is given in Table 2.3. The signs of this correlation matrix give the direction of linear relationship between the corresponding two variables, while the magnitude gives the strength of this correlation. In Chapter 3, we will see that these simple correlations when squared give the percentage of


2.6

Descriptive Statistics

35

9 9

8

LWAGE

LWAGE

8

7

6

7

6

5

5 0

10

20

30

40

50

60

EX

0

4

8

12

16

20

ED

Figure 2.12 Log (Wage) versus Experience

Figure 2.13 Log (Wage) versus Education

9.0 8.5

LWAGE

8.0 7.5 7.0 6.5 6.0 5.5 0

10

20

30

40

50

60

WKS Figure 2.14 Log (Wage) versus Weeks

variation that one of these variables explain in the other. For example, the simple correlation coefficient between log wage and marital status is 0.32. This means that marital status explains (0.32)2 or 10% of the variation in log wage. One cannot emphasize enough how important it is to check one’s data. It is important to compute the descriptive statistics, simple plots of the data and simple correlations. A wrong minimum or maximum could indicate some possible data entry errors. Troughs or peaks in these plots may indicate important events for time series data, like wars or recessions, or influential observations. More on this in Chapter 8. Simple correlation coefficients that equal one indicate perfectly collinear variables and warn of the failure of a linear regression that has both variables included among the regressors, see Chapter 4.


36

Chapter 2: Basic Statistical Concepts

Notes 1. Actually E(s2 ) = σ 2 does not need the normality assumption. This fact along with the proof of (n − 1)s2 /σ 2 ∼ χ2n−1 , under Normality, can be easily shown using matrix algebra and is deferred to Chapter 7. 2. This can be proven using the Chebyshev’s inequality, see Hogg and Craig (1995). 3. See Hogg and Craig (1995) for the type of regularity conditions needed for these distributions.

Problems 1. Variance and Covariance of Linear Combinations of Random Variables. Let a, b, c, d, e and f be arbitrary constants and let X and Y be two random variables. (a) Show that var(a + bX) = b2 var(X). (b) var(a + bX + cY ) = b2 var(X) + c2 var(Y ) + 2bc cov(X, Y ). (c) cov[(a + bX + cY ), (d + eX + f Y )] = be var(X) + cf var(Y ) + (bf + ce) cov(X, Y ). 2. Independence and Simple Correlation. (a) Show that if X and Y are independent, then E(XY ) = E(X)E(Y ) = μx μy where μx = E(X) and μy = E(Y ). Therefore, cov(X, Y ) = E(X − μx )(Y − μy ) = 0.

(b) Show that if Y = a + bX, where a and b are arbitrary constants, then ρxy = 1 if b > 0 and −1 if b < 0.

3. Zero Covariance Does Not Necessarily Imply Independence. Let X = −2, −1, 0, 1, 2 with Pr[X = x] = 1/5. Assume a perfect quadratic relationship between Y and X, namely Y = X 2 . Show that cov(X, Y ) = E(X 3 ) = 0. Deduce that ρXY = correlation (X, Y ) = 0. The simple correlation coefficient ρXY measures the strength of the linear relationship between X and Y . For this example, it is zero even though there is a perfect nonlinear relationship between X and Y . This is also an example of the fact that if ρXY = 0, then X and Y are not necessarily independent. ρxy = 0 is a necessary but not sufficient condition for X and Y to be independent. The converse, however, is true, i.e., if X and Y are independent, then ρXY = 0, see problem 2. 4. The Binomial Distribution is defined as the number of successes in n independent Bernoulli trials with probability of success θ. This discrete probability function is given by   n X f (X; θ) = θ (1 − θ)n−X X = 0, 1, . . . , n X n = n!/[X!(n − X)!]. and zero elsewhere, with X (a) Out of 20 candidates for a job with a probability of hiring of 0.1. Compute the probabilities of getting X = 5 or 6 people hired? n  n = n−X and use that to conclude that b(n, X, ) = b(n, n − X, 1 − θ). (b) Show that X (c) Verify that E(X) = nθ and var(X) = nθ(1 − θ).

(d) For a random sample of size n drawn from the Bernoulli distribution with parameter θ, show ¯ is the MLE of θ. that X ¯ in part (d), is unbiased and consistent for θ. (e) Show that X, ¯ in part (d), is sufficient for θ. (f) Show that X,


Problems

37

¯ in part (d), MVU (g) Derive the Cram´er-Rao lower bound for any unbiased estimator of θ. Is X, for θ? (h) For n = 20, derive the uniformly most powerful critical region of size α ≤ 0.05 for testing H0 ; θ = 0.2 versus H1 ; θ = 0.6. What is the probability of type II error for this test criteria? (i) Form the Likelihood Ratio test for testing H0 ; θ = 0.2 versus H1 ; θ = 0.2. Derive the Wald and LM test statistics for testing H0 versus H1 . When is the Wald statistic greater than the LM statistic? 5. For a random sample of size n drawn from the Normal distribution with mean μ and variance σ 2 . (a) Show that s2 is a sufficient statistic for σ 2 . (b) Using the fact that (n − 1)s2 /σ 2 is χ2n−1 (without proof), verify that E(s2 ) = σ 2 and that var(s2 ) = 2σ 4 /(n − 1) as shown in the text.

(c) Given that σ 2 is unknown, form the Likelihood Ratio test statistic for testing H0 ; μ = 2 versus H1 ; μ = 2. Derive the Wald and Lagrange Multiplier statistics for testing H0 versus H1 . Verify that they are given by the expressions in example 4.

(d) Another derivation of the W ≥ LR ≥ LM inequality for the null hypothesis given in part (c) can be obtained as follows: Let μ , σ 2 be the restricted maximum likelihood estimators under 2 H0 ; μ = μ0 . Let μ , σ  be the corresponding unrestricted maximum likelihood estimators under the alternative H1 ; μ = μ0 . Show that W = −2log[L( μ, σ 2 )/L( μ, σ 2 )] and LM = 2 2 2 μ, σ  )] where L(μ, σ ) denotes the likelihood function. Conclude that W ≥ −2log[L( μ, σ  )/L( LR ≥ LM , see Breusch (1979). This is based on Baltagi (1994). (e) Given that μ is unknown, form the Likelihood Ratio test statistic for testing H0 ; σ = 3 versus H1 ; σ = 3.

(f) Form the Likelihood Ratio test statistic for testing H0 ; μ = 2, σ2 = 4 against the alternative that H1 ; μ = 2 or σ 2 = 4.

(g) For n = 20, s2 = 9 construct a 95% confidence interval for σ 2 .

6. The Poisson distribution can be defined as the limit of a Binomial distribution as n → ∞ and θ → 0 such that nθ = λ is a positive constant. For example, this could be the probability of a rare disease and we are random sampling a large number of inhabitants, or it could be the rare probability of finding oil and n is the large number of drilling sights. This discrete probability function is given by f (X; λ) =

e−λ λX X!

X = 0, 1, 2, . . .

For a random sample from this Poisson distribution (a) Show that E(X) = λ and var(X) = λ. MLE = X. ¯ (b) Show that the MLE of λ is λ

¯ (c) Show that the method of moments estimator of λ is also X. ¯ (d) Show that X is unbiased and consistent for λ. ¯ is sufficient for λ. (e) Show that X ¯ attains (f) Derive the Cram´er-Rao lower bound for any unbiased estimator of λ. Show that X that bound. (g) For n = 9, derive the Uniformly Most Powerful critical region of size α ≤ 0.05 for testing H0 ; λ = 2 versus H1 ; λ = 4.


38

Chapter 2: Basic Statistical Concepts (h) Form the Likelihood Ratio test for testing H0 ; λ = 2 versus H1 ; λ = 2. Derive the Wald and LM statistics for testing H0 versus H1 . When is the Wald test statistic greater than the LM statistic? 7. The Geometric distribution is known as the probability of waiting for the first success in independent repeated trials of a Bernoulli process. This could occur on the 1st, 2nd, 3rd,.. trials. g(X; θ) = θ(1 − θ)X−1 for X = 1, 2, 3, . . . (a) Show that E(X) = 1/θ and var(X) = (1 − θ)/θ2 .

(b) Given a random sample from this Geometric distribution of size n, find the MLE of θ and the method of moments estimator of θ. ¯ is unbiased and consistent for 1/θ. (c) Show that X (d) For n = 20, derive the Uniformly Most Powerful critical region of size α ≤ 0.05 for testing H0 ; θ = 0.5 versus H1 ; θ = 0.3. (e) Form the Likelihood Ratio test for testing H0 ; θ = 0.5 versus H1 ; θ = 0.5. Derive the Wald and LM statistics for testing H0 versus H1 . When is the Wald statistic greater than the LM statistic? 8. The Uniform density, defined over the unit interval [0, 1], assigns a unit probability for all values of X in that interval. It is like a roulette wheel that has an equal chance of stopping anywhere between 0 and 1. f (X) = 1 =0

0≤X ≤1 elsewhere

Computers are equipped with a Uniform (0,1) random number generator so it is important to understand these distributions. (a) Show that E(X) = 1/2 and var(X) = 1/12. (b) What is the Pr[0.1 < X < 0.3]? Does it matter if we ask for the Pr[0.1 ≤ X ≤ 0.3]? 9. The Exponential distribution is given by f (X; θ) =

1 −X/θ e θ

X > 0 and θ > 0

This is a skewed and continuous distribution defined only over the positive quadrant. (a) Show that E(X) = θ and var(X) = θ2 . ¯ (b) Show that  θ MLE = X.

¯ (c) Show that the method of moments estimator of θ is also X. ¯ is an unbiased and consistent estimator of θ. (d) Show that X ¯ is sufficient for θ. (e) Show that X ¯ MVU for θ? (f) Derive the Cram´er-Rao lower bound for any unbiased estimator of θ? Is X (g) For n = 20, derive the Uniformly Most Powerful critical region of size α ≤ 0.05 for testing H0 ; θ = 1 versus H1 ; θ = 2. (h) Form the Likelihood Ratio test for testing H0 ; θ = 1 versus H1 ; θ = 1. Derive the Wald and LM statistics for testing H0 versus H1 . When is the Wald statistic greater than the LM statistic?


Problems

39

10. The Gamma distribution is given by f (X; α, β)

1 X α−1 e−X/β Γ(α)β α =0 =

for X > 0 elsewhere

where α and β > 0 and Γ(α) = (α − 1)! This is a skewed and continuous distribution. (a) Show that E(X) = αβ and var(X) = αβ 2 . (b) For a random sample drawn from this Gamma density, what are the method of moments estimators of α and β? (c) Verify that for α = 1 and β = θ, the Gamma probability density function reverts to the Exponential p.d.f. considered in problem 9. (d) We state without proof that for α = r/2 and β = 2, this Gamma density reduces to a χ2 distribution with r degrees of freedom, denoted by χ2r . Show that E(χ2r ) = r and var(χ2r ) = 2r. (e) For a random sample from the χ2r distribution, show that (X1 X2 , . . . , Xn ) is a sufficient statistic for r. (f) One can show that the square of a N (0, 1) random variable is a χ2 random variable with 1 degree of freedom, see the Appendix to the chapter. Also, one can show that the sum of independent χ2 ’s is a χ2 random variable with degrees of freedom equal the sum of the corresponding degrees of freedom of the individual χ2 ’s, see problem 15. This will prove useful for testing later on. Using these results, verify that the sum of squares of m independent N (0, 1) random variables is a χ2 with m degrees of freedom. 11. The Beta distribution is defined by f (X)

Γ(α + β) α−1 X (1 − X)β−1 Γ(α)Γ(β) =0 =

for 0 < X < 1 elsewhere

where α > 0 and β > 0. This is a skewed continuous distribution. (a) For α = β = 1 this reverts back to the Uniform (0, 1) probability density function. Show that E(X) = (α/α + β) and var(X) = αβ/(α + β)2 (α + β + 1). (b) Suppose that α = 1, find the estimators of β using the method of moments and the method of maximum likelihood. 12. The t-distribution with r degrees of freedom can be defined as the ratio of two independent random variables. The numerator being a N (0, 1) random variable and the denominator being the squareroot of a χ2r random variable divided by its degrees of freedom. The t-distribution is a symmetric distribution like the Normal distribution but with fatter tails. As r → ∞, the t-distribution approaches the Normal distribution. (a) Verify that if X1√ , . . . , Xn are a random sample drawn from a N (μ, σ 2 ) distribution, then ¯ z = (X − μ)/(σ/ n) is N (0, 1).  ¯ − μ)/(s/√n) has a (b) Use the fact that (n − 1)s2 /σ 2 ∼ χ2n−1 to show that t = z/ s2 /σ 2 = (X ¯ t-distribution with (n − 1) degrees of freedom. We use the fact that s2 is independent of X without proving it. (c) For n = 16, x ¯ = 20 and s2 = 4, construct a 95% confidence interval for μ.


40

Chapter 2: Basic Statistical Concepts

13. The F-distribution can be defined as the ratio of two independent χ2 random variables each divided by its corresponding degrees of freedom. It is commonly used to test the equality of variances. Let s21 be the sample variance from a random sample of size n1 drawn from N (μ1 , σ 21 ) and let s22 be the sample variance from another random sample of size n2 drawn from N (μ2 , σ 22 ). We know that (n1 − 1)s21 /σ 21 is χ2(n1 −1) and (n2 − 1)s22 /σ 22 is χ2(n2 −1) . Taking the ratio of those two independent χ2 random variables divided by their appropriate degrees of freedom yields F =

s21 /σ 21 s22 /σ 22

which under the null hypothesis H0 ; σ 21 = σ 22 gives F = s21 /s22 and is distributed as F with (n1 − 1) and (n2 − 1) degrees of freedom. Both s21 and s22 are observable, so F can be computed and compared to critical values for the F -distribution with the appropriate degrees of freedom. Two inspectors drawing two random samples of size 25 and 31 from two shifts of a factory producing steel rods, find that the sampling variance of the lengths of these rods are 15.6 and 18.9 inches squared. Test whether the variances of the two shifts are the same. 14. Moment Generating Function (MGF). (a) Derive the MGF of the Binomial distribution defined in problem 4. Show that it is equal to [(1 − θ) + θet ]n . 1

(b) Derive the MGF of the Normal distribution defined in problem 5. Show that it is eμt+ 2 σ (c) Derive the MGF of the Poisson distribution defined in problem 6. Show that it is e

t

2 2

t

λ(e −1)

.

.

(d) Derive the MGF of the Geometric distribution defined in problem 7. Show that it is θet /[1 − (1 − θ)et ].

(e) Derive the MGF of the Exponential distribution defined in problem 9. Show that it is 1/(1 − θt). (f) Derive the MGF of the Gamma distribution defined in problem 10. Show that it is (1−βt)−α . r Conclude that the MGF of a χ2r is (1 − 2t)− 2 .

(g) Obtain the mean and variance of each distribution by differentiating the corresponding MGF derived in parts (a) through (f). 15. Moment Generating Function Method. . . . , Xn are independent Poisson distributed with parameters (λi ) respec(a) Show that if X1 ,  tively, then Y = ni=1 Xi is Poisson with parameter ni=1 λi .

2 (b) Show that if X1 , . . . , Xn are independent  Normally distributed  with parameters (μi , σ i ), then Y = ni=1 Xi is Normal with mean ni=1 μi and variance ni=1 σ 2i . ¯ ∼ N (μ, σ 2 /n). (c) Deduce from part (b) that if X1 , . . . , Xn are IIN(μ, σ2 ), then X

(d) Show that if X1 , . . . , Xn are independent χ2 distributed n with parameters (ri ) respectively, n then Y = i=1 Xi is χ2 distributed with parameter i=1 ri .

16. Best Linear Prediction. (Problems 16 and 17 are based on Amemiya (1994)). Let X and Y be two random variables with means μX and μY and variances σ 2X and σ 2Y , respectively. Suppose that ρ = correlation(X, Y ) = σ XY /σ X σ Y

where σ XY = cov(X, Y ). Consider the linear relationship Y = α + βX where α and β are scalars: (a) Show that the best linear predictor of Y based on X, where best in this case means the minimum mean squared error predictor which minimizes E(Y − α − βX)2 with respect to α 2  where α   and β is given by Y = α  + βX  = μY − βμ X and β = σ XY /σ X = ρσ Y /σ X .


Problems

41

(b) Show that the var(Y ) = ρ2 σ 2Y and that u  = Y − Y , the prediction error, has mean zero and 2 2 variance equal to (1 − ρ )σ Y . Therefore, ρ2 can be interpreted as the proportion of σ 2Y that is explained by the best linear predictor Y . (c) Show that cov(Y , u ) = 0.

17. The Best Predictor. Let X and Y be the two random variables considered in problem 16. Now consider predicting Y by a general, possibly non-linear, function of X denoted by h(X). (a) Show that the best predictor of Y based on X, where best in this case means the minimum mean squared error predictor that minimizes E[Y − h(X)]2 is given by h(X) = E(Y /X). Hint: Write E[Y − h(X)]2 as E{[Y − E(Y /X)] + [E(Y /X) − h(X)]}2 . Expand the square and show that the cross-product term has zero expectation. Conclude that this mean squared error is minimized at h(X) = E(Y /X). (b) If X and Y are bivariate Normal, show that the best predictor of Y based on X is identical to the best linear predictor of Y based on X. 18. Descriptive Statistics. Using the data used in section 2.6 based on 595 individuals drawn from the Panel Study of Income Dynamics for 1982 and available on the Springer web site as EARN.ASC, replicate the tables and graphs given in that section. More specifically (a) replicate Table 2.1 which gives the descriptive statistics for a subset of the variables in this data set. (b) Replicate Figures 2.6–2.11 which plot the histograms for log wage, weeks worked, education and experience. (c) Replicate Table 2.2 which gives the average log wage for various groups and test the difference between these averages using a t-test. (d) Replicate Figure 2.12 which plots log wage versus experience. Figure 2.13 which plots log wage versus education and Figure 2.14 which plots log wage versus weeks worked. (e) Replicate Table 2.3 which gives the correlation matrix among a subset of these variables. 19. Conflict Among Criteria for Testing Hypotheses: Examples from Non-Normal Distributions. This is based on Baltagi (2000). Berndt and Savin (1977) showed that W ≥ LR ≥ LM for the case of a multivariate regression model with normal distrubances. Ullah and Zinde-Walsh (1984) showed that this inequality is not robust to non-normality of the disturbances. In the spirit of the latter article, this problem considers simple examples from non-normal distributions and illustrates how this conflict among criteria is affected. (a) Consider a random sample x1 , x2 , . . . , xn from a Poisson distribution with parameter λ. Show that for testing λ = 3 versus λ = 3 yields W ≥ LM for x ¯ ≤ 3 and W ≤ LM for x¯ ≥ 3. (b) Consider a random sample x1 , x2 , . . . , xn from an Exponential distribution with parameter θ. Show that for testing θ = 3 versus θ = 3 yields W ≥ LM for 0 < x ¯ ≤ 3 and W ≤ LM for x¯ ≥ 3. (c) Consider a random sample x1 , x2 , . . . , xn from a Bernoulli distribution with parameter θ. Show that for testing θ = 0.5 versus θ = 0.5, we will always get W ≥ LM. Show also, that for testing θ = (2/3) versus θ = (2/3) we get W ≤ LM for (1/3) ≤ x¯ ≤ (2/3) and W ≥ LM for (2/3) ≤ x ¯ ≤ 1 or 0 < x¯ ≤ (1/3).


42

Chapter 2: Basic Statistical Concepts

References More detailed treatment of the material in this chapter may be found in: Amemiya, T. (1994), Introduction to Statistics and Econometrics (Harvard University Press: Cambridge). Baltagi, B.H. (1994), “The Wald, LR, and LM Inequality,” Econometric Theory, Problem 94.1.2, 10: 223-224. Baltagi, B.H. (2000), “Conflict Among Critera for Testing Hypotheses: Examples from Non-Normal Distributions,” Econometric Theory, Problem 00.2.4, 16: 288. Bera A.K. and G. Permaratne (2001), “General Hypothesis Testing,” Chapter 2 in Baltagi, B.H. (ed.), A Companion to Theoretical Econometrics (Blackwell: Massachusetts). Berndt, E.R. and N.E. Savin (1977), “Conflict Among Criteria for Testing Hypotheses in the Multivariate Linear Regression Model,” Econometrica, 45: 1263-1278. Breusch, T.S. (1979), “Conflict Among Criteria for Testing Hypotheses: Extensions and Comments,” Econometrica, 47: 203-207. Buse, A. (1982), “The Likelihood Ratio, Wald, and Lagrange Multiplier Tests: An Expository Note,” The American Statistician, 36 :153-157. DeGroot, M.H. (1986), Probability and Statistics (Addison-Wesley: Mass.). Freedman, D., R. Pisani, R. Purves and A. Adhikari (1991), Statistics (Norton: New York). Freund, J.E. (1992), Mathematical Statistics (Prentice-Hall: New Jersey). Hogg, R.V. and A.T. Craig (1995), Introduction to Mathematical Statistics (Prentice Hall: New Jersey). Jollife, I.T. (1995), “Sample Sizes and the Central Limit Theorem: The Poisson Distribution as an Illustration,” The American Statistician, 49: 269. Kennedy, P. (1992), A Guide to Econometrics (MIT Press: Cambridge). Mood, A.M., F.A. Graybill and D.C. Boes (1974), Introduction to the Theory of Statistics (McGraw-Hill: New York). Spanos, A. (1986), Statistical Foundations of Econometric Modelling (Cambridge University Press: Cambridge). Ullah, A. and V. Zinde-Walsh (1984), “On the Robustness of LM, LR and W Tests in Regression Models,” Econometrica, 52: 1055-1065. Zellner, A. (1971), An Introduction to Bayesian Inference in Econometrics (Wiley: New York).

Appendix Score and Information Matrix: The likelihood function of a sample X1 , . . . , Xn drawn from f (Xi , θ) is really the joint probability density function written as a function of θ: L(θ) = f (X1 , . . . , Xn ; θ)  This probability density function has the property that L(θ)dx = 1 where the integral is over all X1 , . . . , Xn written compactly as one integral over x. Differentiating this multiple integral with respect


Appendix

43

to θ, one gets  ∂L dx = 0 ∂θ Multiplying and dividing by L, one gets       1 ∂L ∂logL L dx = L dx = 0 L ∂θ ∂θ But the score is by definition S(θ) = ∂logL/∂θ. Hence E[S(θ)] = 0. Differentiating again with respect to θ, one gets        2 ∂logL ∂L ∂ logL L+ dx = 0 ∂θ ∂θ ∂θ2 Multiplying and dividing the second term by L one gets   2  ∂logL ∂ 2 logL + =0 E ∂θ ∂θ2 or

∂ 2 logL E − ∂θ2

=E



∂logL ∂θ

2 

= E[S(θ)]2

But var[S(θ)] = E[S(θ)]2 since E[S(θ)] = 0. Hence I(θ) = var[S(θ)]. Moment Generating Function (MGF): For the random variable X, the expected value of a special function of X, namely eXt is denoted by MX (t) = E(eXt ) = E(1 + Xt + X 2

t2 t3 + X 3 + ..) 2! 3!

where the second equality follows from the Taylor series expansion of eXt around zero. Therefore, MX (t) = 1 + E(X)t + E(X 2 )

t3 t2 + E(X 3 ) + .. 2! 3!

This function of t generates the moments of X as coefficients of an infinite polynomial in t. For example, μ = E(X) = coefficient of t, and E(X 2 )/2 is the coefficient of t2 , etc. Alternatively, one can differentiate ′ (0), i.e., the first derivative of MX (t) with this MGF with respect to t and obtain μ = E(X) = MX r r respect to t evaluated at t = 0. Similarly, E(X ) = MX (0) which is the r-th derivative of MX (t) with respect to t evaluated at t = 0. For example, for the Bernoulli distribution; 1 MX (t) = E(eXt ) = X=0 eXt θX (1 − θ)1−X = θet + (1 − θ)

′′ ′′ ′ ′ so that MX (0) = θ. (t) = θet which means that E(X 2 ) = MX (0) = θ = E(X) and MX (t) = θet and MX Hence,

var(X) = E(X 2 ) − (E(X))2 = θ − θ2 = θ(1 − θ). For the Normal distribution, see problem 14, it is easy to show that if X ∼ N (μ, σ 2 ), then MX (t) = 1 2 2 ′ ′′ eμt+ 2 σ t and MX (0) = E(X) = μ and MX (0) = E(X 2 ) = σ 2 + μ2 . There is a one-to-one correspondence between the MGF when it exists and the corresponding p.d.f. 2 This means that if Y has a MGF given by e2t+4t then Y is normally distributed with mean 2 and variance 8. Similarly, if Z has a MGF given by (et + 1)/2, then Z is Bernoulli distributed with θ = 1/2.


44

Chapter 2: Basic Statistical Concepts

Change of Variable: If X ∼ N (0, 1), then one can find the distribution function of Y = |X| by using the Distribution Function method. By definition the distribution function of y is defined as G(y)

= Pr[Y ≤ y] = Pr[|X| ≤ y] = Pr[−y ≤ X ≤ y] = Pr[X ≤ y] − Pr[X ≤ −y] = F (y) − F (−y)

so that the distribution function of Y, G(y), can be obtained from the distribution function of X, F (x). Since the N (0, 1) distribution is symmetric around zero, then F (−y) = 1 − F (y) and substituting that in G(y) we get G(y) = 2F (y) − 1. Recall, that the p.d.f. of Y is given by g(y) = G′ (y). Hence, g(y) = f (y) + f (−y) and this reduces to 2f (y) if the distribution is symmetric around zero. So that if √ √ 2 2 f (x) = e−x /2 / 2π for −∞ < x < +∞ then g(y) = 2f (y) = 2e−y /2 / 2π for y ≥ 0. Let us now find the distribution of Z = X 2 , the square of a N (0, 1) random variable. Note that dZ/dX = 2X which is positive when X > 0 and negative when X < 0. The change of variable method cannot be applied since Z = X 2 is not a monotonic transformation over the entire domain of X. However, using Y = |X|, we get Z = Y 2 = (|X|)2 and dZ/dY = 2Y which is always non-negative since Y is nonnegative. In this case, the change of variable method√states that the p.d.f. of Z is obtained from that of Y by substituting the inverse transformation Y = Z into the p.d.f. of Y and multiplying it by the absolute value of the derivative of the inverse transformation: √ 2 1 dY 1 | = √ e−z/2 | √ | = √ z −1/2 e−z/2 for z ≥ 0 h(z) = g( z) · | dZ 2 z 2π 2π It is clear why this transformation will not work for X since Z = X 2 has two √ solutions for the inverse √ transformation, X = ± Z, whereas, there is one unique solution for Y = Z since it is non-negative. Using the results of problem 10, one can deduce that Z has a gamma distribution with α = 1/2 and β = 2. This special Gamma density function is a χ2 distribution with 1 degree of freedom. Hence, we have shown that the square of a N (0, 1) random variable has a χ21 distribution. n Finally, if X1 , . . . , Xn are independently distributed then the distribution function of Y = i=1 Xi can be obtained from that of the Xi ’s using the Moment Generating Function (MGF) method: n

= E(eY t ) = E[e( i=1 Xi )t ] = E(eX1 t )E(eX2 t )..E(eXn t ) = MX1 (t)MX2 (t)..MXn (t)

MY (t)

If in addition these Xi ’s are identically distributed, then MXi (t) = MX (t) for i = 1, . . . , n and MY (t) = [MX (t)]n t For example, if X1 , . . . , X n are IID Bernoulli (θ), then MXi (t) = MX (t) = θe + (1 − θ) for i = 1, . . . , n.  Hence the MGF of Y = ni=1 Xi is given by

MY (t) = [MX (t)]n = [θet + (1 − θ)]n

This can be easily shown to be the MGF of the Binomial distribution given in problem 14. This proves that the sum of n independent and identically distributed Bernoulli random variables with parameter θ is a Binomial random variable with same parameter θ. ¯ −μ X √ Central Limit Theorem: If X1 , . . . , Xn are IID(μ, σ2 ) from an unknown distribution, then Z = σ/ n is asymptotically distributed as N (0, 1). Proof : We assume that the MGF of the Xi ’s exist and derive the MGF of Z. Next, we show that lim 2 MZ (t) as n → ∞ is e1/2t which is the MGF of N (0, 1) distribution. First, note that Z=

n

Xi − nμ Y − nμ √ √ = σ n σ n

i=1


Appendix where Y =

45

n

MZ (t)

Xi with MY (t) = [MX (t)]n . Therefore,   â&#x2C6;&#x161;  â&#x2C6;&#x161; â&#x2C6;&#x161;  = E(eZt ) = E e(Y tâ&#x2C6;&#x2019;nÎźt)/Ď&#x192; n = eâ&#x2C6;&#x2019;nÎźt/Ď&#x192; n E eY t/Ď&#x192; n â&#x2C6;&#x161; â&#x2C6;&#x161; â&#x2C6;&#x161; â&#x2C6;&#x161; = eâ&#x2C6;&#x2019;nÎźt/Ď&#x192; n MY (t/Ď&#x192; n) = eâ&#x2C6;&#x2019;nÎźt/Ď&#x192; n [MX (t/Ď&#x192; n)]n

i=1

Taking log of both sides we get logMZ (t) =

t t2 t3 â&#x2C6;&#x2019;nÎźt â&#x2C6;&#x161; + nlog[1 + â&#x2C6;&#x161; E(X) + 2 E(X 2 ) + 3 â&#x2C6;&#x161; E(X 3 ) + ..] Ď&#x192; n Ď&#x192; n 2Ď&#x192; n 6Ď&#x192; n n

s2 s3 Using the Taylor series expansion log(1 + s) = s â&#x2C6;&#x2019; + â&#x2C6;&#x2019; .. we get 2 3 

â&#x2C6;&#x161; t t2 nÎź t3 2 3 logMZ (t) = â&#x2C6;&#x2019; t + n Îź â&#x2C6;&#x161; + 2 E(X ) + 3 â&#x2C6;&#x161; E(X ) + .. Ď&#x192; Ď&#x192; n 2Ď&#x192; n 6Ď&#x192; n n 2

2 3 t 1 t t 2 3 â&#x2C6;&#x161; â&#x2C6;&#x161; + E(X ) + 3 E(X ) + .. â&#x2C6;&#x2019; Îź 2 Ď&#x192; n 2Ď&#x192; 2 n 6Ď&#x192; n n 

3 1 t2 t3 t 2 3 + Îź â&#x2C6;&#x161; + 2 â&#x2C6;&#x161; E(X ) + 3 â&#x2C6;&#x161; E(X ) + .. â&#x2C6;&#x2019; .. 3 Ď&#x192; n 2Ď&#x192; n 6Ď&#x192; n n Collecting powers of t, we get   â&#x2C6;&#x161;  â&#x2C6;&#x161;  (E(X 2 ) nÎź nÎź Îź2 + â&#x2C6;&#x2019; t2 log MZ (t) = â&#x2C6;&#x2019; t+ Ď&#x192; Ď&#x192; 2Ď&#x192; 2 2Ď&#x192; 2   1 2ÎźE(X 2 ) 1 Îź3 E(X 3 ) â&#x2C6;&#x161; â&#x2C6;&#x161; â&#x2C6;&#x161; + â&#x2C6;&#x2019; ¡ + t3 + .. 6Ď&#x192; 3 n 2 2Ď&#x192; 3 n 3 Ď&#x192;3 n Therefore



 3 E(X 3 ) ÎźE(X 2 ) Îź3 t â&#x2C6;&#x161; + .. â&#x2C6;&#x2019; + 6 2 3 Ď&#x192;3 n â&#x2C6;&#x161; note that the coeďŹ&#x192;cient of t3 is 1/ n times a constant. Therefore, this coeďŹ&#x192;cient goes to zero as n â&#x2020;&#x2019; â&#x2C6;&#x17E;. â&#x2C6;&#x161; Similarly, it can be shown that the coeďŹ&#x192;cient of tr is 1/ nrâ&#x2C6;&#x2019;2 times a constant for r â&#x2030;Ľ 3. Hence, logMZ (t) =

1 2 t + 2

lim logMZ (t) =

nâ&#x2020;&#x2019;â&#x2C6;&#x17E;

1 2 t 2

and

1 2

lim MZ (t) = e 2 t

nâ&#x2020;&#x2019;â&#x2C6;&#x17E;

which is the MGF of a standard normal distribution. The Central Limit Theorem is a powerful tool for asymptotic inference. In real life we do not know what distribution we are sampling from, but as long as the sample drawn is random and we average (or sum) and standardize then as n â&#x2020;&#x2019; â&#x2C6;&#x17E;, the resulting standardized statistic has an asymptotic N (0, 1) distribution that can be used for inference. Using a random number generator from say the uniform distribution on the computer, one can generate samples of size n = 20, 30, 50 from this distribution and show how the sampling distribution of the sum (or average) when it is standardized closely approximates the N (0, 1) distribution. The real question for the applied researcher is how large n should be to invoke the Central Limit Theorem. This depends on the distribution we are drawing from. For a Bernoulli distribution, a larger n is needed the more asymmetric this distribution is i.e., if θ = 0.1 rather than 0.5. In fact, Figure 2.15 shows the Poisson distribution with mean = 15. This looks like a good approximation for a Normal distribution even though it is a discrete probability function. Problem 15 shows that the sum of n independent identically distributed Poisson random variables with parameter Îť is a Poisson random variable with parameter (nÎť). This means that if Îť = 0.15, an n of 100 will lead to the distribution of the sum being Poisson (nÎť = 15) and the Central Limit Theorem seems well approximated.


46

Chapter 2: Basic Statistical Concepts

p (x) 0.12 0.10 0.08 0.06 0.04 0.02 0.00 0

5

10

15

20

25

30

x

Figure 2.15 Poisson Probability Distribution, Mean = 15 p (x ) 0.4

0.3

0.2

0.1

0

0 2 4 6 Figure 2.16 Poisson Probability Distribution, Mean = 1.5

8

10

x

However, if λ = 0.015, an n of 100 will lead to the distribution of the sum being Poisson (nλ = 1.5) which is given in Figure 2.16. This Poisson probability function is skewed and discrete and does not approximate well a normal density. This shows that one has to be careful in concluding that n = 100 is a large enough sample for the Central Limit Theorem to apply. We showed in this simple example that this depends on the distribution we are sampling from. This is true for Poisson (λ = 0.15) but not Poisson (λ = 0.015), see Joliffe (1995). The same idea can be illustrated with a skewed Bernoulli distribution. Conditional Mean and Variance: Two random variables X and Y are bivariate Normal if they have the following joint distribution:      x − μX 2 y − μY 2 1 1  exp − + f (x, y) = 2(1 − ρ2 ) σX σY 2πσ X σ Y 1 − ρ2     y − μY x − μX −2ρ σX σY


Appendix

47

where â&#x2C6;&#x2019;â&#x2C6;&#x17E; < x < +â&#x2C6;&#x17E;, â&#x2C6;&#x2019;â&#x2C6;&#x17E; < y < +â&#x2C6;&#x17E;, E(X) = ÎźX , E(Y ) = ÎźY , var(X) = Ď&#x192; 2X , var(Y ) = Ď&#x192; 2Y and Ď = correlation (X, Y ) = cov(X, Y )/Ď&#x192; X Ď&#x192; Y . This joint density can be rewritten as  2 

1 Ď&#x192;Y 1  exp â&#x2C6;&#x2019; 2 f (x, y) = â&#x2C6;&#x161; (x â&#x2C6;&#x2019; ÎźX ) y â&#x2C6;&#x2019; ÎźY â&#x2C6;&#x2019; Ď 2Ď&#x192; Y (1 â&#x2C6;&#x2019; Ď 2 ) Ď&#x192;X 2Ď&#x20AC;Ď&#x192; Y 1â&#x2C6;&#x2019; Ď 2  1 1 exp â&#x2C6;&#x2019; 2 (x â&#x2C6;&#x2019; ÎźX )2 = f (y/x)f1 (x) ¡â&#x2C6;&#x161; 2Ď&#x192;X 2Ď&#x20AC;Ď&#x192; X where f1 (x) is the marginal density of X and f (y/x) is the conditional density of Y given X. In this Ď&#x192; case, X â&#x2C6;ź N (ÎźX , Ď&#x192; 2X ) and Y /X is Normal with mean E(Y /X) = ÎźY + Ď Y (x â&#x2C6;&#x2019; Îźx ) and variance given Ď&#x192;X by var(Y /X) = Ď&#x192; 2Y (1 â&#x2C6;&#x2019; Ď 2 ). By symmetry, the roles of X and Y can be interchanged and one can write f (x, y) = f (x/y) f2 (y) where f2 (y) is the marginal density of Y . In this case, Y â&#x2C6;ź N (ÎźY , Ď&#x192; 2Y ) and X/Y is Normal with mean Ď&#x192; E(X/Y ) = ÎźX + Ď X (y â&#x2C6;&#x2019; ÎźY ) and variance given by var(X/Y ) = Ď&#x192; 2X (1 â&#x2C6;&#x2019; Ď 2 ). If Ď = 0, then f (y/x) = Ď&#x192;Y f2 (y) and f (x, y) = f1 (x)f2 (y) proving that X and Y are independent. Therefore, if cov(X, Y ) = 0 and X and Y are bivariate Normal, then X and Y are independent. In general, cov(X, Y ) = 0 alone does not necessarily imply independence, see problem 3. One important and useful property is the law of iterated expectations. This says that the expectation of any function of X and Y say h(X, Y ) can be obtained as follows: E[h(X, Y )] = EX EY /X [h(X, Y )] where the subscript Y /X on E means the conditional expectation of Y given that X is treated as a constant. The next expectation EX treats X as a random variable. The proof is simple.  +â&#x2C6;&#x17E;  +â&#x2C6;&#x17E; h(x, y)f (x, y)dxdy E[h(X, Y )] = â&#x2C6;&#x2019;â&#x2C6;&#x17E;

â&#x2C6;&#x2019;â&#x2C6;&#x17E;

where f (x, y) is the joint density of X  and Y . But f (x, y) can be written as f (y/x)f1 (x), hence  +â&#x2C6;&#x17E;  +â&#x2C6;&#x17E; â&#x2C6;&#x2019;â&#x2C6;&#x17E; â&#x2C6;&#x2019;â&#x2C6;&#x17E; h(x, y)f (y/x)dy f1 (x)dx = EX EY /X [h(X, Y )].

E[h(X, Y )] =

Example: This law of iterated expectation can be used to show that for the bivariate Normal density, the parameter Ď is indeed the correlation coeďŹ&#x192;cient of X and Y . In fact, let h(X, Y ) = XY , then Ď&#x192; E(XY ) = EX EY /X (XY /X) = EX XE(Y /X) = EX X[ÎźY + Ď Y (X â&#x2C6;&#x2019; ÎźX )] Ď&#x192;X Ď&#x192; = ÎźX ÎźY + Ď Y Ď&#x192; 2X = ÎźX ÎźY + Ď Ď&#x192; Y Ď&#x192; X Ď&#x192;X

Rearranging terms, one gets Ď = [E(XY ) â&#x2C6;&#x2019; ÎźX ÎźY ]/Ď&#x192; X Ď&#x192; Y = Ď&#x192; XY /Ď&#x192; X Ď&#x192; Y as required. Another useful result pertains to the unconditional variance of h(X, Y ) being the sum of the mean of the conditional variance and the variance of the conditional mean: var(h(X, Y )) = EX varY /X [h(X, Y )] + varX EY /X [h(X, Y )] Proof : We will write h(X, Y ) as h to simplify the presentation varY /X (h) = EY /X (h2 ) â&#x2C6;&#x2019; [EY /X (h)]2 and taking expectations with respect to X yields EX varY /X (h) = EX EY /X (h2 ) â&#x2C6;&#x2019; EX [EY /X (h)]2 = E(h2 ) â&#x2C6;&#x2019; EX [EY /X (h)]2 . Also, varX EY /X (h) = EX [EY /X (h)]2 â&#x2C6;&#x2019; (EX [EY /X (h)])2 = EX [EY /X (h)]2 â&#x2C6;&#x2019; [E(h)]2 adding these two terms yields E(h2 ) â&#x2C6;&#x2019; [E(h)]2 = var(h).


CHAPTER 3

Simple Linear Regression 3.1

Introduction

In this chapter, we study extensively the estimation of a linear relationship between two variables, Yi and Xi , of the form: Yi = α + βXi + ui

i = 1, 2, . . . , n

(3.1)

where Yi denotes the i-th observation on the dependent variable Y which could be consumption, investment or output, and Xi denotes the i-th observation on the independent variable X which could be disposable income, the interest rate or an input. These observations could be collected on firms or households at a given point in time, in which case we call the data a cross-section. Alternatively, these observations may be collected over time for a specific industry or country in which case we call the data a time-series. n is the number of observations, which could be the number of firms or households in a cross-section, or the number of years if the observations are collected annually. α and β are the intercept and slope of this simple linear relationship between Y and X. They are assumed to be unknown parameters to be estimated from the data. A plot of the data, i.e., Y versus X would be very illustrative showing what type of relationship exists empirically between these two variables. For example, if Y is consumption and X is disposable income then we would expect a positive relationship between these variables and the data may look like Figure 3.1 when plotted for a random sample of households. If α and β were known, one could draw the straight line (α + βX) as shown in Figure 3.1. It is clear that not all the observations (Xi , Yi ) lie on the straight line (α + βX). In fact, equation (3.1) states that the difference between each Yi and the corresponding (α + βXi ) is due to a random error ui . This error may be due to (i) the omission of relevant factors that could influence consumption, other than disposable income, like real wealth or varying tastes, or unforseen events that induce households to consume more or less, (ii) measurement error, which could be the result of households not reporting their consumption or income accurately, or (iii) wrong choice of a linear relationship between consumption and income, when the true relationship may be nonlinear. These different causes of the error term will have different effects on the distribution of this error. In what follows, we consider only disturbances that satisfy some restrictive assumptions. In later chapters we relax these assumptions to account for more general kinds of error terms. In real life, α and β are not known, and have to be estimated from the observed data {(Xi , Yi ) for i = 1, 2, . . . , n}. This also means that the true line (α + βX) as well as the true disturbances (the ui ’s) are unobservable. In this case, α and β could be estimated by the best fitting line through the data. Different researchers may draw different lines through the same data. What makes one line better than another? One measure of misfit is the amount of error from the  i , where the hat (ˆ) denotes observed Yi to the guessed line, let us call the latter Yi = α  + βX a guess on the appropriate parameter or variable. Each observation (Xi , Yi ) will have a corresponding observable error attached to it, which we will call ei = Yi − Yi , see Figure 3.2. In other words, we obtain the guessed Yi , (Yi ) corresponding to each Xi from the guessed line,


50

Chapter 3: Simple Linear Regression

 i . Next, we find our error in guessing that Yi , by subtracting the actual Yi from the α  + βX guessed Yi . The only difference between Figure 3.1 and Figure 3.2 is the fact that Figure 3.1 draws the true consumption line which is unknown to the researcher, whereas Figure 3.2 is a guessed consumption line drawn through the data. Therefore, while the ui ’s are unobservable, the ei ’s are observable. Note that there will be n errors for each line, one error corresponding to every observation. Similarly, there will be another set of n errors for another guessed line drawn through the data. For each guessed line, we can summarize its corresponding errors by one number, the sum of squares of these errors, which seems to be a natural criterion for penalizing a wrong guess. Note that a simple sum of these errors is not a good choice for a measure of misfit since positive errors end up canceling negative errors when both should be counted in our measure. However, this does not mean that the sum of squared error is the only single measure of misfit. Other measures include the sum of absolute errors, but this latter measure is mathematically more difficult to handle. Once the measure of misfit is chosen, α and β could then be estimated by minimizing this measure. In fact, this is the idea behind least squares estimation.

3.2

Least Squares Estimation and the Classical Assumptions

Least squares minimizes the residual sum of squares where the residuals are given by  i ei = Yi − α  − βX

i = 1, 2, . . . , n

 denote guesses on the regression parameters α and β, respectively. The residual and α  and β    i )2 is minimized by the two  − βX sum of squares denoted by RSS = ni=1 e2i = ni=1 (Yi − α first-order conditions:   n  n Xi = 0 ∂( ni=1 e2i )/∂α = −2 ni=1 ei = 0; or α−β (3.2) i=1 Yi − n i=1 n  n    n X 2 = 0  ∂( ni=1 e2i )/∂β = −2 ni=1 ei Xi = 0; or Xi − β i=1 Yi Xi − α i=1 i

(3.3)

i=1

Solving the least squares normal equations given in (3.2) and (3.3) for α and β one gets  OLS X  OLS = n xi yi / n x2 ¯ and β (3.4) α  OLS = Y¯ − β i=1 i=1 i  ¯ = n Xi /n, yi = Yi − Y¯ , xi = Xi − X, ¯ n x2 = n X 2 − nX ¯ 2, where Y¯ =  ni=1 Yi /n, X i=1 i=1 i i=1 i    n n n n 2 2 2 ¯ ¯ ¯ i=1 yi = i=1 Yi − nY and i=1 xi yi = i=1 Xi Yi − nX Y . These estimators are subscripted by OLS denoting the ordinary least squares estimators. The  OLS residuals ei = Yi − α  OLS − β OLS Xi automatically satisfy the  two numerical relationships given by (3.2) and (3.3). The first relationship states that (i) ni=1 ei = 0, the residuals sum to zero. This is true as long as there is a constant in the regression. This numerical property of the least squares residuals also implies that the estimated regression line passes through the ¯ Y¯ ). To see this, average the residuals, or equation (3.2), this gives immediately sample means (X, n  ¯ Y¯ = α  OLS + β OLS X. The second relationship states that (ii) i=1 ei Xi = 0, the residuals and the explanatory variable are uncorrelated. Other numerical properties that the OLS estimators n  n n  satisfy are the following: (iii) i=1 Yi = i=1 Yi and (iv) i=1 ei Yi = 0. Property (iii) states that the sum of the estimated Yi ’s or the predicted Yi ’s from the sample is equal to the sum of the


3.2

Least Squares Estimation and the Classical Assumptions

Y

Y

ui

= + >X i

+ + + + + + + + + +

Consumption

+ + + + + + + + + + + +

+ + + + ++ + + + + +

=ˆ + >ˆX i

= + >X i

Yi

ei

=ˆ + >ˆX i

Disposable Income

Xi

Figure 3.1 ‘True’ Consumption Function

+ + + + + + + + + +

+ +

+ + +

Consumption

Yi

51

X

+ + + + + + + + + +

+ + + + ++ + + + + +

+ + +

Disposable Income

Xi

X

Figure 3.2 Estimated Consumption Function

actual Yi ’s. Property (iv) states that the OLS residuals and the predicted Yi ’s are uncorrelated. The proof of (iii) and (iv) follow from (i) and (ii) see problem 1. Of course, underlying our estimation of (3.1) is the assumption that (3.1) is the true model generating the data. In this case, (3.1) is linear in the parameters α and β, and contains only one explanatory variable Xi besides the constant. The inclusion of other explanatory variables in the model will be considered in Chapter 4, and the relaxation of the linearity assumption will be considered in Chapters 8 and 13. In order to study the statistical properties of the OLS estimators of α and β, we need to impose some statistical assumptions on the model generating the data. Assumption 1: The disturbances have zero mean, i.e., E(ui ) = 0 for every i = 1, 2, . . . , n. This assumption is needed to insure that on the average we are on the true line. To see what happens if E(ui ) = 0, consider the case where households consistently under-report their consumption by a constant amount of δ dollars, while their income is measured accurately, say by cross-referencing it with their IRS tax forms. In this case, (Observed Consumption) = (T rue Consumption) − δ and our regression equation is really (T rue Consumption)i = α + β(Income)i + ui But we observe, (Observed Consumption)i = α + β(Income)i + ui − δ This can be thought of as the old regression equation with a new disturbance term u∗i = ui − δ. Using the fact that δ > 0 and E(ui ) = 0, one gets E(u∗i ) = −δ < 0. This says that for all households with the same income, say $20, 000, their observed consumption will be on the average below that predicted from the true line [α+β($20, 000)] by an amount δ. Fortunately, one


52

Chapter 3: Simple Linear Regression

can deal with this problem of constant but non-zero mean of the disturbances by reparametizing the model as (Observed Consumption)i = α∗ + β(Income)i + ui where α∗ = α − δ. In this case, E(ui ) = 0 and α∗ and β can be estimated from the regression. Note that while α∗ is estimable, α and δ are non-estimable. Also note that for all $20, 000 income households, their average consumption is [(α − δ) + β($20, 000)].

Assumption 2: The disturbances have a constant variance, i.e., var(ui ) = σ 2 for every i = 1, 2, . . . , n. This insures that every observation is equally reliable. To see what this assumption means, consider the case where var(ui ) = σ 2i , for i = 1, 2, . . . , n. In this case, each observation has a different variance. An observation with a large variance is less reliable than one with a smaller variance. But, how can this differing variance happen? In the case of consumption, households with large disposable income (a large Xi , say $100, 000) may be able to save more (or borrow more to spend more) than households with smaller income (a small Xi , say $10, 000). In this case, the variation in consumption for the $100, 000 income household will be much larger than that for the $10, 000 income household. Therefore, the corresponding variance for the $100, 000 observation will be larger than that for the $10, 000 observation. Consequences of different variances for different observations will be studied more rigorously in Chapter 5. Assumption 3: The disturbances are not correlated, i.e., E(ui uj ) = 0 for i = j, i, j = 1, 2, . . . , n. Knowing the i-th disturbance does not tell us anything about the j-th disturbance, for i = j. For the consumption example, the unforseen disturbance which caused the i-th household to consume more, (like a visit of a relative), has nothing to do with the unforseen disturbances of any other household. This is likely to hold for a random sample of households. However, it is less likely to hold for a time series study of consumption for the aggregate economy, where a disturbance in 1945, a war year, is likely to affect consumption for several years after that. In this case, we say that the disturbance in 1945 is related to the disturbances in 1946, 1947, and so on. Consequences of correlated disturbances will be studied in Chapter 5. Assumption 4: The explanatory variable X is nonstochastic, i.e., fixed in repeated samples,  and hence, not correlated with the disturbances. Also, ni=1 x2i /n = 0 and has a finite limit as n tends to infinity.

This assumption states that we have at least two distinct values for X. This makes sense, since ¯ = X, the common we need at least two distinct points to draw a straight line. Otherwise X n 2 ¯ value, and x = X − X = 0, which violates i=1 xi = 0. In practice, one always has several distinct values of X. More importantly, this assumption implies that X is not a random variable and hence is not correlated with the disturbances. In section 5.3, we will relax the assumption of a non-stochastic X. Basically, X becomes a random variable and our assumptions have to be recast conditional on the set of X’s that are observed. This is the more realistic case with economic data. The zero mean assumption becomes E(ui /X) = 0, the constant variance assumption becomes var(ui /X) = σ 2 , the no serial correlation assumption becomes E(ui uj /X) = 0 for i = j. The conditional expectation here is with respect to every observation on Xi from i = 1, 2, . . . n. Of course, one can show that if E(ui /X) = 0 for all i, then Xi and ui are not correlated. The reverse is not necessarily true, see


3.2

Least Squares Estimation and the Classical Assumptions

53

Consumption

Y + +++ + ++ + + + + ++ + + + ++

a + bX

+ + + + + + + + + + + ++ ++ + ++ + + + + +

mx

Disposable Income

X

Figure 3.3 Consumption Function with Cov(X, u) > 0

problem 3 of Chapter 2. That problem shows that two random variables, say ui and Xi could be uncorrelated, i.e., not linearly related when in fact they are nonlinearly related with ui = Xi2 . Hence, E(ui /Xi ) = 0 is a stronger assumption than ui and Xi are not correlated. By the law of iterated expectations given in the Appendix of Chapter 2, E(ui /X) = 0 implies that E(ui ) = 0. It also implies that ui is uncorrelated with any function of Xi . This is a stronger assumption than ui is uncorrelated with Xi . Therefore, conditional on Xi , the mean of the disturbances is zero and does not depend on Xi . In this case, E(Yi /Xi ) = α + βXi is linear in α and β and is assumed to be the true conditional mean of Y given X. To see what a violation of assumption 4 means, suppose that X is a random variable and that X and u are positively correlated, then in the consumption example, households with income above the average income will be associated with disturbances above their mean of zero, and hence positive disturbances. Similarly, households with income below the average income will be associated with disturbances below their mean of zero, and hence negative disturbances. This means that the disturbances are systematically affected by values of the explanatory variable and the scatter of the data will look like Figure 3.3. Note that if we now erase the true line (α + βX), and estimate this line from the data, the least squares line drawn through the data is going to have a smaller intercept and a larger slope than those of the true line. The scatter should look like Figure 3.4 where the disturbances are random variables, not correlated with the Xi ’s, drawn from a distribution with zero mean and constant variance. Assumptions 1 and 4 insure that E(Yi /Xi ) = α + βXi , i.e., on the average we are on the true line. Several economic models will be studied where X and u are correlated. The consequences of this correlation will be studied in Chapters 5 and 11. We now generate a data set which satisfies all four classical assumptions. Let α and β take the arbitrary values, say 10 and 0.5 respectively, and consider a set of 20 fixed X’s, say income classes from $10 to $105 (in thousands of dollars), in increments of $5, i.e., $10, $15, $20, $25, . . . , $105. Our consumption variable Yi is constructed as (10 + 0.5Xi + ui ) where ui is a


54

Chapter 3: Simple Linear Regression

f (u )

Y

= + >X

X1 X2 . . .

Xn

X Figure 3.4 Random Disturbances around the Regression

disturbance which is a random draw from a distribution with zero mean and constant variance, say σ 2 = 9. Computers generate random numbers with various distributions. In this case, Figure 3.4 would depict our data, with the true line being (10 + 0.5X) and ui being random draws from the computer which are by construction independent and identically distributed with mean zero and variance 9. For every set of 20 ui ’s randomly generated, given the fixed Xi ’s, we obtain a corresponding set of 20 Yi ’s from our linear regression model. This is what we mean in assumption 4 when we say that the X’s are fixed in repeated samples. Monte Carlo experiments generate a large number of samples, say a 1000, in the fashion described above. For each data set generated, least squares can be performed and the properties of the resulting estimators which are derived analytically in the remainder of this chapter, can be verified. For example, the average of the 1000 estimates of α and β can be compared to their true values to see whether these least squares estimates are unbiased. Note what will happen to Figure 3.4 if E(ui ) = −δ where δ > 0, or var(ui ) = σ 2i for i = 1, 2, . . . , n. In the first case, the mean of f (u), the probability density function of u, will shift off the true line (10+ 0.5X) by −δ. In other words, we can think of the distributions of the ui ’s, shown in Figure 3.4 , being centered on a new imaginary line parallel to the true line but lower by a distance δ. This means that one is more likely to draw negative disturbances than positive disturbances, and the observed Yi ’s are more likely to be below the true line than above it. In the second case, each f (ui ) will have a different variance, hence the spread of this probability density function will vary with each observation. In this case, Figure 3.4 will have a distribution for the ui ’s which has a different spread for each observation. In other words, if the ui ’s are say normally distributed, then u1 is drawn from a N (0, σ 21 ) distribution, whereas u2 is drawn from a N (0, σ 22 ) distribution, and so on. Violation of the classical assumptions can also be studied using Monte Carlo experiments, see Chapter 5.


3.3

3.3

Statistical Properties of Least Squares

55

Statistical Properties of Least Squares

(i) Unbiasedness  Given assumptions 1-4, it is easy to show that β OLS is unbiased for β. In fact, using equation (3.4) one can write OLS = n xi yi / n x2 = n xi Yi / n x2 = β + n xi ui / n x2 (3.5) β i=1 i=1 i i=1 i=1 i i=1 i=1 i   where the second equality follows from the fact that yi = Yi − Y¯ and ni=1 xi Y¯ = Y¯ ni=1 xi = 0. The third equality follows from substituting Yi from (3.1) and using the fact that ni=1 xi = 0. Taking expectations of both sides of (3.5) and using assumptions 1 and 4, one can show that  OLS ) = β. Furthermore, one can derive the variance of β OLS from (3.5) since E(β   n n 2 2 2   (3.6) var(β OLS ) = E(β OLS − β) = E( i=1 xi ui / i=1 xi ) n n 2 2 n = var( i=1 xi ui / i=1 xi ) = σ / i=1 x2i

where the last equality uses assumptions 2 and 3, i.e., that the ui ’s are not correlated with each other and that their variance is constant, see problem 4. Note that the variance of the OLS estimator of β depends upon σ 2 , the variance of the disturbances  in the true model, and on the variation in X. The larger the variation in X the larger is ni=1 x2i and the smaller is the OLS . variance of β (ii) Consistency

  Next, we show that β OLS is consistent for β. A sufficient condition for consistency is that β OLS  OLS is unbiased and its variance tends to zero as n tends to infinity. We have already shown β to be unbiased, it remains to show that its variance tends to zero as n tends to infinity. n 2 2  lim var(β OLS ) = lim [(σ /n)/( i=1 xi /n)] = 0 n→∞

n→∞

 where the second equality follows from the fact that (σ 2 /n) → 0 and ( ni=1 x2i /n) = 0 and has   a finite limit, see assumption 4. Hence, plim β OLS = β and β OLS is consistent  for β. Similarly  one can show that α  OLS is unbiased and consistent for α with variance σ 2 ni=1 Xi2 /n ni=1 x2i ,  n 2  ¯ 2 and cov( αOLS , β OLS ) = −Xσ / i=1 xi , see problem 5. (iii) Best Linear Unbiased

 OLS OLS as n wi Yi where wi = xi / n x2 . This proves that β Using (3.5) one can write β i=1 i=1 i is a linear combination of the Yi ’s, with weights wi satisfying the following properties: n n n n 2 2 (3.7) i=1 wi = 0; i=1 wi Xi = 1; i=1 wi = 1/ i=1 xi

 The next theorem shows that among all linear unbiased estimators of β, it is β OLS which has the smallest variance. This is known as the Gauss-Markov Theorem.  = n ai Yi for β, where the ai ’s denote Theorem 1: Consider any arbitrary linear estimator β i=1  is unbiased for β, and assumptions 1 to 4 are satisfied, then var(β)  ≥ arbitrary constants. If β  var(β ). OLS


56

Chapter 3: Simple Linear Regression

 one gets β  = α n ai +β n ai Xi + n ai ui . For β  Proof: Substituting Yi from (3.1) into β, i=1  i=1 i=1  n n  =α to be unbiased for β it must followthat E(β) i=1 ai +β i=1 ai Xi = β for all observations  i = 1, 2, . . . , n. This means that ni=1 ai = 0 and ni=1 ai Xi = 1 for all i = 1, 2, . . . , n. Hence,  = β + n ai ui with var(β)  = var(n ai ui ) = σ 2 n a2 where the last equality follows β i=1 i=1 i=1 i from assumptions 2 and 3. But the ai ’s are constants which differ from the wi ’s, the weights of the OLS estimator, by some other constants, say di ’s, i.e., ai = wi + di for i = 1, 2, . . . , n. Using n the properties of the a ’s and w one can deduce similar properties on the d ’s i.e., i i i i=1 di = 0 n and i=1 di Xi = 0. In fact, n n n n 2 2 2 i=1 ai = i=1 di + i=1 wi + 2 i=1 wi di n n n 2 where i=1 wi d i = i=1 xi di / i=1 xi = 0. This follows from the definition of wi and the fact n n that i=1 di = i=1 di Xi = 0. Hence,  = σ2 var(β)

n

2 i=1 ai

= σ2

n

2 i=1 di

+ σ2

n

2 i=1 wi

2  = var(β OLS ) + σ

n

2 i=1 di

  ≥ var(β  Since σ 2 ni=1 d2i is non-negative, this proves that var(β) OLS ) with the equality holding  reduces to β  only if di = 0 for all i = 1, 2, . . . , n, i.e., only if ai = wi , in which case β OLS .  Therefore, any linear estimator of β, like β that is unbiased for β has variance at least as large  OLS ). This proves that β OLS is BLUE, Best among all Linear Unbiased Estimators of as var(β β. Similarly, one can show that α  OLS is linear in Yi and has the smallest variance among all linear unbiased estimators of α, if assumptions 1 to 4 are satisfied, see problem 6. This result implies that the OLS estimator of α is also BLUE.

3.4

Estimation of σ 2

The variance of the regression disturbances σ 2 is unknown and has to be estimated. In fact,  both the variance of β  OLS depend upon σ 2 , see (3.6) and problem 5. An OLS and that n of 2α 2 2 unbiased estimator for σ is s = i=1 ei /(n − 2). To prove this, we need the fact that

   ei = Yi − α  OLS − β ¯) OLS Xi = yi − β OLS xi = (β − β OLS )xi + (ui − u n  ¯ where u ¯ =  OLS = Y¯ − β OLS X and the third i=1 ui /n. The second equality substitutes α equality substitutes yi = βxi + (ui − u ¯). Hence,  n n n 2 n 2 2   ¯)2 − 2(β ¯), OLS − β) i=1 xi + i=1 (ui − u i=1 xi (ui − u i=1 ei = (β OLS − β) and

n n  2 2 2 n 2  E( ni=1 e2i ) = i=1 xi var(β OLS ) + (n − 1)σ − 2E( i=1 xi ui ) / i=1 xi

= σ 2 + (n − 1)σ 2 − 2σ 2 = (n − 2)σ 2   where the first equality uses the fact that E( ni=1 (ui − u ¯)2 ) = (n − 1)σ 2 and β OLS − β = n n  n 2 2 2  i=1 xi ui / i=1 xi . The second equality uses the fact that var(β OLS ) = σ / i=1 xi and   E( ni=1 xi ui )2 = σ 2 ni=1 x2i .


3.5

Maximum Likelihood Estimation

57

 Therefore, E(s2 ) = E( ni=1 e2i /(n − 2)) = σ 2 .  Intuitively, the estimator of σ 2 could be obtained from ni=1 (ui − u ¯)2 /(n − 1) if the true disturbances were known.  Since the u’s are not known, consistent estimates of them are used.  These are the ei ’s. Since ni=1 ei = 0, our estimator of σ 2 becomes ni=1 e2i /(n − 1). Taking expectations we find that the correct divisor ought to be (n−2) and not (n−1) for this estimator to be unbiased for σ 2 . This is plausible, since we have estimated two parameters α and β in obtaining the ei ’s, and there are only n − 2 independent pieces of information left in the data. To prove this fact, consider the OLS normal equations given in (3.2) and (3.3). These equations represent two relationships involving the ei ’s. Therefore, knowing (n − 2) of the ei ’s we can deduce the remaining two ei ’s from (3.2) and (3.3).

3.5

Maximum Likelihood Estimation

Assumption 5: The ui ’s are independent and identically distributed N (0, σ 2 ). This assumption allows us to derive distributions of estimators and other test statistics. In  fact using (3.5) one can easily see that β OLS is a linear combination of the ui ’s. But, a linear combination of normal random variables see Chapter 2,    is itself a normal random variable,  problem 15. Hence, β is N (β, σ 2 / n x2 ). Similarly α  OLS is N (α, σ 2 n X 2 /n n x2 ), OLS

i=1

i

i=1

i

i=1

i

and Yi is N (α+βXi , σ 2 ). Moreover, we can write the density function of the u’s njoint2 probability 2 2 2 n/2 as f (u1 , u2 , . . . , un ; α, β, σ ) = (1/2πσ ) exp(− i=1 ui /2σ ). To get the likelihood function we make the transformation ui = Yi − α− βXi and note that the Jacobian of the transformation is 1. Therefore, f (Y1 , Y2 , . . . , Yn ; α, β, σ 2 ) = (1/2πσ 2 )n/2 exp{− Taking the log of this likelihood, we get logL(α, β, σ 2 ) = −(n/2)log(2πσ 2 ) −

n

i=1 (Yi

n

i=1 (Yi

− α − βXi )2 /2σ 2 }

− α − βXi )2 /2σ 2

(3.8)

(3.9)

Maximizing this likelihood with respect to α, β and σ 2 one gets the maximum likelihood estimators (MLE). However, only the second term in the log likelihood contains α and β and that term (without the negative sign) has already been minimized with respect to α and β in (3.2)   and (3.3) giving us the OLS estimators. Hence, α  M LE = α  OLS and β M LE = β OLS . Similarly, 2 and setting this derivative equal to zero one gets by differentiating logL with respect to σ  σ 2M LE = ni=1 e2i /n, see problem 7. Note that this differs from s2 only in the divisor. In fact, E( σ 2M LE ) = (n − 2)σ 2 /n = σ 2 . Hence, σ 2M LE is biased but note that it is still asymptotically unbiased. So far, the gains from imposing assumption 5 are the following: The likelihood can be formed, maximum likelihood estimators can be derived, and distributions can be obtained for these estimators. One can also derive the Cram´er-Rao lower bound for unbiased estimators of the 2  parameters and show that the α  OLS and β OLS attain this bound whereas s does not. This derivation is postponed until Chapter 7. In fact, one can show following the theory of complete 2  sufficient statistics that α  OLS , β OLS and s are minimum variance unbiased estimators for α, β 2  and σ , see Chapter 2. This is a stronger result (for α  OLS and β OLS ) than that obtained using the Gauss-Markov Theorem. It says, that among all unbiased estimators of α and β, the OLS


58

Chapter 3: Simple Linear Regression

estimators are the best. In other words, our set of estimators include now all unbiased estimators and not just linear unbiased estimators. This stronger result is obtained at the expense of a stronger distributional assumption, i.e., normality. If the distribution of the disturbances is not normal, then OLS is no longer MLE. In this case, MLE will be more efficient than OLS as long as the distribution of the disturbances is correctly specified. Some of the advantages and disadvantages of MLE were discussed in Chapter 2. 2  We found the distributions of α  OLS , β OLS , now we give that of s . In Chapter 7, it is shown n 2 2 that i=1 ei /σ is a chi-squared with (n − 2) degrees of freedom. Also, s2 is independent of  α  OLS and β OLS . This is useful for test of hypotheses. In fact, the major gain from assumption 5 is that we can perform test of hypotheses. OLS , one gets z = (β  OLS − β)/(σ 2 / n x2 ) 12 ∼ Standardizing the normal random variable β i=1 i N (0, 1). Also, (n − 2)s2 /σ 2 is distributed as χ2n−2 . Hence, one can divide z, a N (0, 1) random variable, by the square root of (n − 2)s2 /σ 2 divided by its degrees of freedom (n − 2) to  get a t-statistic with (n − 2) degrees of freedom. The resulting statistic is tobs = (β OLS − n 1 2 2 β)/(s / i=1 xi ) 2 ∼ tn−2 , see problem 8. This statistic can be used to test H0 ; β = β 0 , versus H1 ; β = β 0 , where β 0 is a known constant. Under H0 , tobs can be calculated and its value can be compared to a critical value from a t-distribution with (n − 2) degrees of freedom, at a specified critical value of α%. Of specific interest is the hypothesis H0 ; β = 0, which states that there is no linear relationship between Yi and Xi . Under H0 ,  OLS /(s2 / tobs = β

n

2 12 i=1 xi )

 OLS /se(  OLS ) =β  β

n 2 2 12  where se(  β OLS ) = (s / i=1 xi ) . If |tobs | > tα/2;n−2 , then H0 is rejected at the α% significance level. tα/2;n−2 represents a critical value obtained from a t-distribution with n − 2 degrees of freedom. It is determined such that the area to its right under a tn−2 distribution is equal to α/2. Similarly one can get a confidence interval for β by using the fact that, Pr[−tα/2;n−2 < tobs <   tα/2;n−2 ] = 1 − α and substituting for tobs its value derived above as (β  β OLS − β)/se( OLS ).   Since the critical values are known, β and se(  β ) can be calculated from the data, the OLS OLS following (1 − α)% confidence interval for β emerges  OLS ). OLS ± tα/2;n−2 se(  β β

Tests of hypotheses and confidence intervals on α and σ 2 can be similarly constructed using the normal distribution of α  OLS and the χ2n−2 distribution of (n − 2)s2 /σ 2 .

3.6

A Measure of Fit

We have obtained the least squares estimates of α, β and σ 2 and found their distributions under normality of the disturbances. We have also learned how to test hypotheses regarding these parameters. Now we turn to a measure of fit for this estimated regression line. Recall, that ei = Yi − Yi where Yi denotes the predicted Yi from n the least squares regression nline at the nvalue   Xi , i.e., α  OLS + β OLS Xi . Using the fact that i=1 ei = 0, we deduce that i=1 Yi = i=1 Yi , and therefore, Y¯ = Y . The actual and predicted values of Y have the same sample mean, see numerical properties (i) and (iii) of the OLS estimators discussed in section 2. This is true


3.6

A Measure of Fit

59

as long as there is a constant in the regression. Adding and subtracting Y¯ from ei , we get ei = yi − yi , or yi = ei + yi . Squaring and summing both sides: n

2 i=1 yi

=

n

2 i=1 ei

+

n

i2 i=1 y

+2

n

i i=1 ei y

=

n

2 i=1 ei

+

n

i2 i=1 y

OLS xi and where the last equality follows from the fact that yi = β n

i i=1 ei y

=

n

 =0

i=1 ei Yi

n

i=1 ei xi

(3.10)

= 0. In fact,

means that the OLS residuals are uncorrelated with the predicted values from the regression, see numerical properties (ii) and (iv) of the OLS estimates discussed in section  3.2. In other words, (3.10) says that the total variation in Yi , around its sample mean Y¯ i.e., ni=1 yi2 , can be  2 n x2 , decomposed into two parts: the first is the regression of squares ni=1 yi2 = β OLS i=1 i n sums and the second is the residual sums of squares i=1 e2i . In fact, regressing Y on a constant yields α  OLS = Y¯ , see problem 2, and the unexplained residual sums of squares of this naive model is n

i=1 (Yi

−α  OLS )2 =

n

i=1 (Yi

− Y¯ )2 =

n

2 i=1 yi .

 Therefore, ni=1 yi2 in (3.10) gives the explanatory power of X after the constant is fit. Using this decomposition, one can define the explanatory power of the regression as the ratio of regression sums of squares to the total sums of squares. In other words, define the n n 2 = 2/ 2 and this value is clearly between 0 and 1. In fact, dividing (3.10) by y  y R i i i=1 i=1 n 2 n 2 n n 2 2 2 i=1 ei is a measure of misfit which was i=1 yi . The i=1 ei / i=1 yi one gets R = 1 −  minimized by least squares. If ni=1 e2i is large, this means that the regression is not explaining  a lot of the variation in Y and hence, the R2 value would be small. Alternatively, if the ni=1 e2i is small, then the fit is good and R2 is large. In fact, for a perfect where all the observations n fit, 2 = 0 and R2 = 1. The other e lie on the fitted line, Yi = Yi and ei = 0, which means that i=1 i  extreme case is where the regression sums of squares ni=1 yi2  = 0. In other the linear n words, n 2 2 regression explains nothing of the variation in Yi . In this case, i=1 yi = i=1 ei and R2 = 0.  Note that since ni=1 yi2 = 0 implies yi = 0 for every i, which in turn means that Yi = Y¯ for every i. The fitted regression line is a horizontal line drawn at Y = Y¯ , and the independent variable X does not have any explanatory power in a linear relationship with Y . Note that R2 has two alternative meanings: (i) It is the simple squared correlation coefficient between Yi and Yi , see problem 9. Also, for the simple regression case, (ii) it is the simple squared correlation between X and Y . This means that before one runs the regression of Y on 2 which in turn tells us the proportion of the variation in Y that will X, one can compute rxy be explained by X. If this number is pretty low, we have a weak linear relationship between Y and X and we know that a poor fit will result if Y is regressed on X. It is worth emphasizing that R2 is a measure of linear association between Y and X. There could exist, for example, a perfect quadratic relationship between X and Y , yet the estimated least squares line through the data is a flat line implying that R2 = 0, see problem 3 of Chapter 2. One should also be suspicious of least squares regressions with R2 that are too close to 1. In some cases, we may not want to include a constant in the regression. In such cases, one should use an uncentered R2 as a measure fit. The appendix to this chapter defines both centered and uncentered R2 and explains the difference between them.


60

Chapter 3: Simple Linear Regression

3.7

Prediction

Let us now predict Y0 given X0 . Usually this is done for a time series regression, where the researcher is interested in predicting the future, say one period ahead. This new observation Y0 is generated by (3.1), i.e., Y0 = α + βX0 + u0

(3.11)

What is the Best Linear Unbiased Predictor (BLUP) of E(Y0 )? From (3.11), E(Y0 ) = α + βX0  is a linear combination of α and β. Using the Gauss-Markov result, Y0 = α  OLS + β OLS X0 is  2 2 ¯ BLUE for α+βX0 and the variance of this predictor of E(Y0 ) is σ [(1/n)+(X0 − X) / ni=1 x2i ], see problem 10. But, what if we are interested in the BLUP for Y0 itself ? Y0 differs from E(Y0 ) by u0 , and the best predictor of u0 is zero, so the BLUP for Y0 is still Y0 . The forecast error is Y0 − Y0 = [Y0 − E(Y0 )] + [E(Y0 ) − Y0 ] = u0 + [E(Y0 ) − Y0 ]

where u0 is the error committed even if the true regression line is known, and E(Y0 ) − Y0 is the difference between the sample and population regression lines. Hence, the variance of the forecast error becomes: ¯ 2 / n x2 ] var(u0 ) + var[E(Y0 ) − Y0 ] + 2cov[u0 , E(Y0 ) − Y0 ] = σ 2 [1 + (1/n) + (X0 − X) i=1 i

This says that the variance of the forecast error is equal to the variance of the predictor of E(Y0 ) plus the var(u0 ) plus twice the covariance of the predictor of E(Y0 ) and u0 . But, this last covariance is zero, since u0 is a new disturbance and is not correlated with the disturbances in the sample upon which Yi is based. Therefore, the predictor of the average consumption of a $20, 000 income household is the same as the predictor of consumption for a specific household whose income is $20, 000. The difference is not in the predictor itself but in the variance attached to it. The latter variance being larger only by σ 2 , the variance of u0 . The variance of the predictor therefore, depends upon σ 2 , the sample size, the variation in the X’s, and how far X0 is from To summarize, the smaller σ 2 is, the larger n nthe 2sample mean of the observed data. ¯ and i=1 xi are, and the closer X0 is to X, the smaller is the variance of the predictor. One can construct 95% confidence intervals to these predictionsfor every value of X0 . In fact, this n 2 12  ¯ 2 is ( αOLS + β OLS X0 ) ± t.025;n−2 {s[1 + (1/n) + (X0 − X) / i=1 xi ] } where s replaces σ, and t.025;n−2 represents the 2.5% critical value obtained from a t-distribution with n − 2 degrees of freedom. Figure 3.5 shows this confidence band around the estimated regression line. This is a ¯ as expected, and widens as we predict away from X. ¯ hyperbola which is the narrowest at X

3.8

Residual Analysis

A plot of the residuals of the regression is very important. The residuals are consistent estimates of the true disturbances. But unlike the ui ’s, these ei ’s are not independent. In fact, the OLS normal equations (3.2)and (3.3) give us two relationships between these residuals. Therefore, knowing (n − 2) of these residuals the remaining two residuals can be deduced. If we had the true ui ’s, and we plotted them, they should look like a random scatter around the horizontal axis with no specific pattern to them. A plot of the ei ’s that shows a certain pattern like a set of positive residuals followed by a set of negative residuals as shown in Figure 3.6(a) may be


3.8

Residual Analysis

61

Y + + + + + + ++ +

Y

+

Yˆi = =ˆ + >ˆX i

+ +

+ +

+ +

+ + +

+ ++ + +

+

X

X Figure 3.5 95% Confidence Bands

indicative of a violation of one of the 5 assumptions imposed on the model, or simply indicating a wrong functional form. For example, if assumption 3 is violated, so that the ui ’s are say positively correlated, then it is likely to have a positive residual followed by a positive one, and a negative residual followed by a negative one, as observed in Figure 3.6(b). Alternatively, if we fit a linear regression line to a true quadratic relation between Y and X, then a scatter of residuals like that in Figure 3.6(c) will be generated. We will study how to deal with this violation and how to test for it in Chapter 5. Large residuals are indicative of bad predictions in the sample. A large residual could be a typo, where the researcher entered this observation wrongly. Alternatively, it could be an influential observation, or an outlier which behaves differently from the other data points in the sample and therefore, is further away from the estimated regression line than the other data points. The fact that OLS minimizes the sum of squares of these residuals means that a large weight is put on this observation and hence it is influential. In other words, removing this observation from the sample may change the estimates and the regression line significantly. For more on the study of influential observations, see Belsely, Kuh and Welsch (1980). We will focus on this issue in Chapter 8 of this book.

Yi

ei

Yˆi = =ˆ + >ˆX i

Yi = f ( X i ) Yi

+

Yˆi = =ˆ + >ˆX i

_ +

0

-

+

+

> Xi

>

(a) Figure 3.6 Positively Correlated Residuals

(b)

Xi

>

(c)

Xi


62

Chapter 3: Simple Linear Regression

ei

+ + + + ++ + + + + ++ + + + + + +++ + + + + + + ++ + + + + + + + + + + + + + + ++ + + +++ + + + + ++ + + + ++ + + ++ +

0

> Xi

Figure 3.7 Residual Variation Growing with X

One can also plot the residuals versus the Xi ’s. If a pattern like Figure 3.7 emerges, this could be indicative of a violation of assumption 2 because the variation of the residuals is growing with Xi when it should be constant for all observations. Alternatively, it could imply a relationship between the Xi ’s and the true disturbances which is a violation of assumption 4. In summary, one should always plot the residuals to check the data, identify influential observations, and check violations of the 5 assumptions underlying the regression model. In the next few chapters, we will study various tests of the violation of the classical assumptions. Most of these tests are based on the residuals of the model. These tests along with residual plots should help the researcher gauge the adequacy of his or her model.

Table 3.1 Simple Regression Computations OBS

Consumption yi

Income xi

yi=Yi−Y¯

1 2 3 4 5 6 7 8 9 10

4.6 3.6 4.6 6.6 7.6 5.6 5.6 8.6 8.6 9.6

5 4 6 8 8 7 6 9 10 12

–1.9 –2.9 –1.9 0.1 1.1 –0.9 –0.9 2.1 2.1 3.1

SUM

6.5

75

0

MEAN

6.5

7.5

xi yi

x2i

Yˆi

ei

–2.5 –3.5 –1.5 0.5 0.5 –0.5 –1.5 1.5 2.5 4.5

4.75 10.15 2.85 0.05 0.55 0.45 1.35 3.15 5.25 13.95

6.25 12.25 2.25 0.25 0.25 0.25 2.25 2.25 6.25 20.25

4.476190 3.666667 5.285714 6.904762 6.904762 6.095238 5.285714 7.714286 8.523810 10.142857

0.123810 –0.066667 –0.685714 –0.304762 0.695238 –0.495238 0.314286 0.885714 0.076190 –0.542857

0

42.5

52.5

65

0

¯ xi =Xi-X

6.5


3.9

3.9

Numerical Example

63

Numerical Example

Table 3.1 gives the annual consumption of 10 households each selected randomly from a group of households with a fixed personal disposable income. Both income and consumption are measured in $10, 000, so that the first household earns $50, 000 and consumes $46, 000 annually. It is worthwhile doing the computations necessary to obtain the least squares regression estimates of consumption on income in this simple case and to compare them with those obtained from ¯ = 7.5 and form two a regression package. In order to do this, we first compute Y¯ = 6.5 and X OLS we need n xi yi , ¯ To get β new columns of data made up of yi = Yi − Y¯ and xi = Xi − X. i=1 so we multiply these last two columns by each other and sum to get 42.5. The denominator of n 2 2  β OLS is given by i=1 xi . This is why we square the xi column to get xi and sum to obtain  52.5. Our estimate of β OLS = 42.5/52.5 = 0.8095 which is the estimated marginal propensity to consume. This is the extra consumption brought about by an extra dollar of disposable income.  OLS X ¯ = 6.5 − (0.8095)(7.5) = 0.4286 α  OLS = Y¯ − β

This is the estimated consumption at zero personal disposable income. The fitted values or  predicted values from this regression are computed from Yi = α  OLS + β OLS Xi = 0.4286 +  0.8095Xi and are given in Table 3.1. Note that the mean of Yi is equal to the mean of Yi confirming one of the numerical from  properties of least squares. The residuals are computed  ei = Yi − Yi and they satisfy ni=1 ei = 0. It is left to the reader to verify that ni=1 ei Xi = 0. The residual sum  of squares is obtained by squaring thecolumn of residuals and summing it. This gives us ni=1 e2i = 2.495238. This means that s2 = ni=1 e2i /(n − 2) = 0.311905. Its square root is given by s = 0.558. Thisis known as the standard error of the regression. In this case, n 2 2  the estimated var(β OLS ) is s / i=1 xi = 0.311905/52.5 = 0.005941 and the estimated

¯2 X (7.5)2 1 1 = 0.365374 + n + = 0.311905 var( α ) = s2 2 n 10 52.5 i=1 xi

Taking the square root of these estimated variances, we get the estimated standard errors of   α  OLS and β  αOLS ) = 0.60446 and se(  β OLS given by se( OLS ) = 0.077078. Since the disturbances are normal, the OLS estimators are also the maximum likelihood estimators, and are normally distributed themselves. For the null hypothesis H0a ; β = 0; the observed t-statistic is  OLS − 0)/se( OLS ) = 0.809524/0.077078 = 10.50 tobs = (β  β

and this is highly significant, since Pr[|t8 | > 10.5] < 0.0001. This probability can be obtained using most regression packages. It is also known as the p-value or probability value. It shows that this t-value is highly unlikely and we reject Ha0 that β = 0. Similarly, the null hypothesis H0b ; α = 0, gives an observed t-statistic of tobs = ( αOLS − 0)/se(  αOLS ) = 0.428571/0.604462 = 0.709, which is not significant, since its p-value is Pr[|t8 | > 0.709] < 0.498. Hence, we do not reject the null hypothesis H0b that  α = 0. The total sum of squares is ni=1 yi2 = ni=1 (Yi −Y¯ )2 which can be obtained by squaring the yi column in Table 3.1 and summing. This yields ni=1 yi2 = 36.9. Also, the regression sum of   squares = ni=1 yi2 = ni=1 (Yi − Y¯ )2 which can be obtained by subtracting Y¯ = Y = 6.5 from the Yi column, squaring that column and summing. This yields 34.404762. This could have also been obtained as


64

Chapter 3: Simple Linear Regression

2 n x2 = (0.809524)2 (52.5) = 34.404762. =β OLS i=1 i   n = ni=1 yi2 − ni=1 e2i =36.9 − 2.495238 = 34.404762 as required. A final check is that i=1 yi2 2 = ( n x y )2 /( n x2 )( n y 2 ) = (42.5)2 /(52.5)(36.9) = 0.9324. Recall, that R2 = rxy i=1 i i i=1 i  i=1 i  This could have also been obtained as R2 = 1 − ( ni=1 e2i / ni=1 yi2 ) = 1 − (2.495238/36.9) = 0.9324, or as   R2 = ry2yˆ = ni=1 yi2 / ni=1 yi2 = 34.404762/36.9 = 0.9324. n

i2 i=1 y

This means that personal disposable income explains 93.24% of the variation in consumption. A plot of the actual, predicted and residual values versus time is given in Figure 3.8. This was done using EViews. 12

8

R esidual A ctual F itted

4

0

1

2

3

4

5

6

7

8

9

10

Figure 3.8 Residual Plot

3.10

Empirical Example

Table 3.2 gives (i) the logarithm of cigarette consumption (in packs) per person of smoking age (> 16 years) for 46 states in 1992, (ii) the logarithm of real price of cigarettes in each state, and (iii) the logarithm of real disposable income per capita in each state. This is drawn from Baltagi and Levin (1992) study on dynamic demand for cigarettes. It can be downloaded as Cigarett.dat from the Springer web site.


3.10

Empirical Example

Table 3.2 Cigarette Consumption Data LNC: log of consumption (in packs) per person of smoking age (>16) LNP: log of real price (1983$/pack) LNY: log of real disposable income per-capita (in thousand 1983$) OBS

STATE

LNC

LNP

LNY

1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 24 25 26 27 28 29 30 31 32 33 34 35 36 37 38 39 40 41 42 43 44 45 46

AL AZ AR CA CT DE DC FL GA ID IL IN IA KS KY LA ME MD MA MI MN MS MO MT NE NV NH NJ NM NY ND OH OK PA RI SC SD TN TX UT VT VA WA WV WI WY

4.96213 4.66312 5.10709 4.50449 4.66983 5.04705 4.65637 4.80081 4.97974 4.74902 4.81445 5.11129 4.80857 4.79263 5.37906 4.98602 4.98722 4.77751 4.73877 4.94744 4.69589 4.93990 5.06430 4.73313 4.77558 4.96642 5.10990 4.70633 4.58107 4.66496 4.58237 4.97952 4.72720 4.80363 4.84693 5.07801 4.81545 5.04939 4.65398 4.40859 5.08799 4.930651 4.66134 4.82454 4.83026 5.00087

0.20487 0.16640 0.23406 0.36399 0.32149 0.21929 0.28946 0.28733 0.12826 0.17541 0.24806 0.08992 0.24081 0.21642 â&#x20AC;&#x201C;0.03260 0.23856 0.29106 0.12575 0.22613 0.23067 0.34297 0.13638 0.08731 0.15303 0.18907 0.32304 0.15852 0.30901 0.16458 0.34701 0.18197 0.12889 0.19554 0.22784 0.30324 0.07944 0.13139 0.15547 0.28196 0.19260 0.18018 0.11818 0.35053 0.12008 0.22954 0.10029

4.64039 4.68389 4.59435 4.88147 5.09472 4.87087 5.05960 4.81155 4.73299 4.64307 4.90387 4.72916 4.74211 4.79613 4.64937 4.61461 4.75501 4.94692 4.99998 4.80620 4.81207 4.52938 4.78189 4.70417 4.79671 4.83816 5.00319 5.10268 4.58202 4.96075 4.69163 4.75875 4.62730 4.83516 4.84670 4.62549 4.67747 4.72525 4.73437 4.55586 4.77578 4.85490 4.85645 4.56859 4.75826 4.71169

Data: Cigarette Consumption of 46 States in 1992

65


66

Chapter 3: Simple Linear Regression

Table 3.3 Cigarette Consumption Regression Analysis of Variance Source Model Error

DF 1 44

Root MSE Dep Mean C.V.

Sum of Squares 0.48048 1.16905

Mean Square 0.48048 0.02657

F Value 18.084

0.16300 4.84784 3.36234

R-square Adj R-sq

0.2913 0.2752

Prob > F 0.0001

Parameter Estimates Variable INTERCEP LNP

DF 1 1

Parameter Estimate 5.094108 –1.198316

Standard Error 0.06269897 0.28178857

T for H0: Parameter=0 81.247 –4.253

Prob > |T| 0.0001 0.0001

.4

Residuals

.2 .0 –.2 –.4 –.6 –.1

.0

.1

.2

.3

.4

Log of Real Price (1983$/Pack) Figure 3.9 Residuals versus LNP

Table 3.3 gives the SAS output for the regression of logC on logP . The price elasticity of demand for cigarettes in this simple model is (dlogC/logP ) which is the slope coefficient. This is estimated to be −1.198 with a standard error of 0.282. This says that a 10% increase in real price of cigarettes has an estimated 12% drop in per capita consumption of cigarettes. The R2 of this regression is 0.29, s2 is given by the Mean Square Error of the regression which is 0.0266. Figure 3.9 plots the residuals of this regression versus the independent variable, while Figure 3.10 plots the predictions along with the 95% confidence interval band for these predictions. One observation clearly stands out as an influential observation given its distance from the rest of the data and that is the observation for Kentucky, a producer state with very low real price. This observation almost anchors the straight line fit through the data. More on influential observations in Chapter 8.


Problems

67

Predicted Value of LNC

5.6

5.2

4.8

4.4

4.0 –.1

.0

.1

.2

.3

.4

Log of Real Price (1983$/Pack) Figure 3.10 95% Confidence Band for Predicted Values

Problems 1. For the simple regression with a constant Yi = α + βXi + ui , given in equation (3.1) verify the following numerical properties of the OLS estimators: n n n n n   i=1 ei = 0, i=1 ei Xi = 0, i=1 ei Yi = 0, i=1 Yi = i=1 Yi

2. For the regression with only a constant Yi = α + ui with ui ∼ IID(0, σ 2 ), show that the least squares estimate  is α  OLS = Y¯ , var( αOLS ) = σ 2 /n, and the residual sums of squares is n of α  n 2 2 ¯ i=1 (Yi − Y ) . i=1 yi = 3. For the simple regression without a constant Yi = βXi + ui , with ui ∼ IID(0, σ 2 ). (a) Derive the OLS estimator of β and find its variance.

(b) What numerical properties of the OLS estimators described in problem 1 still hold for this model? (c) derive the maximum likelihood estimator of β and σ 2 under the assumption ui ∼ IIN(0, σ 2 ). (d) Assume σ 2 is known. Derive the Wald, LM and LR tests for H0 ; β = 1 versus H1 ; β = 1.    4. Use the fact that E( ni=1 xi ui )2 = ni=1 nj=1 xi xj E(ui uj ); and assumptions 2 and 3 to prove equation (3.6). 5. Using the regression given in equation (3.1):  ¯ ¯; and deduce that E( (a) Show that α  OLS = α + (β − β αOLS ) = α. OLS )X + u   n n 2  (b) Using the fact that β OLS − β = i=1 xi ; use the  results in part (a) to show that  i=1 xi ui /  ¯ 2 / n x2 )] = σ 2 n X 2 /n n x2 . var( αOLS ) = σ 2 [(1/n) + (X i=1 i i=1 i i=1 i

(c) Show that α  OLS is consistent for α. n 2 ¯ 2   ¯ (d) Show that cov( αOLS , β OLS ) = −Xvar(β OLS ) = −σ X/ i=1 xi . This means that the sign ¯ If X ¯ is positive, this covariance will be of the covariance is determined by the sign of X.  negative. This also means that if α  OLS is over-estimated, β OLS will be under-estimated.


68

Chapter 3: Simple Linear Regression 6. Using the regression given in equation (3.1): n ¯ i and wi = xi / n x2 . (a) Prove that α  OLS = i=1 λi Yi where λi = (1/n) − Xw i=1 i n n (b) Show that i=1 λi = 1 and i=1 λi Xi = 0. n n (c) Prove  = i=1 bi Yi must satisfy i=1 bi = 1 and n that any other linear estimator of α, say α  to be unbiased for α. i=1 bi Xi = 0 for α n n (d) Let bi = λi + fi ; show that i=1 fi = 0 and i=1 fi Xi = 0. n n   n n αOLS ) + σ 2 i=1 fi2 . (e) Prove that var( α) = σ 2 i=1 b2i = σ 2 i=1 λ2i + σ 2 i=1 fi2 =var( 7.

  (a) Differentiate (3.9) with respect to α and β and show that α  MLE = α  OLS , β MLE = β OLS .  n 2 (b) Differentiate (3.9) with respect to σ 2 and show that σ MLE = i=1 e2i /n.

8. It is well known that a standard normal random variable N (0, 1) divided by a square root of a chi1 squared random variable divided by its degrees of freedom (χ2ν /ν) 2 results in a random variable that is t-distributed with ν degrees of freedom, provided the N (0, 1) and the χ2 variables are n 2 12  independent, see Chapter 2. Use this fact to show that (β OLS − β)/[s/( i=1 xi ) ] ∼ tn−2 . n n n n 2   9. (a) Using the fact that R2 = i=1 yi2 / i=1 yi2 ; yi = β OLS xi ; and β OLS = i=1 xi , i=1 xi yi / 2 2 show that R = rxy where,    2 = ( ni=1 xi yi )2 /( ni=1 x2i )( ni=1 yi2 ). rxy n n (b) Using the fact that y i +  ei , show that i=1 yi yi = i=1 yi2 , and hence, deduce that i = y  ry2yˆ = ( ni=1 yi yi )2 /( ni=1 yi2 )( ni=1 yi2 ) is equal to R2 .

10. Prediction. Consider the problem of predicting Y0 from (3.11). Given X0 , (a) Show that E(Y0 ) = α + βX0 . (b) Show that Y0 is unbiased for E(Y0 ).

   (c) Show that var(Y0 ) = var( αOLS )+ X02 var(β αOLS , β OLS )+ 2X0 cov( OLS ). Deduce that var(Y0 ) ¯ 2 / n x2 ]. = σ 2 [(1/n) + (X0 − X) i=1 i n n (d)  Consider a linear predictor of E(Y0 ), say Y0 = i=1 ai Yi , show that i=1 ai = 1 and n a X = X for this predictor to be unbiased for E(Y ). i i 0 0 i=1 n n 2 (e) Show that the var(Y0 ) = σ 2 i=1 a2i . Minimize i=1 ai subject to the restrictions given    OLS + β in (d). Prove that the resulting predictor is Y0 = α OLS X0 and that the minimum n 2 2 2 ¯ variance is σ [(1/n) + (X0 − X) / i=1 xi ].

11. Optimal Weighting of Unbiased Estimators. This is based on Baltagi (1995). For the simple regression without a constant Yi = βXi + ui , i = 1, 2, . . . , N ; where β is a scalar and ui ∼ IID(0, σ 2 ) independent of Xi . Consider the following three unbiased estimators of β:  = n Xi Yi / n X 2 , β  = Y¯ /X ¯ β i 2 1 i=1

i=1

and

 = n (Xi − X)(Y ¯ 2, ¯ i − Y¯ )/ n (Xi − X) β 3 i=1 i=1 ¯ = n Xi /n and Y¯ = n Yi /n. where X i=1 i=1

 ,β  ) = var(β  ) > 0, and that ρ = (the correlation coefficient of β  and (a) Show that cov(β 1 2 1 12 1 1     β 2 ) = [var(β 1 )/var(β 2 )] 2 with 0 < ρ12 ≤ 1. Show that the optimal combination of β 1 and  , given by β  = αβ  + (1 − α)β  where −∞ < α < ∞ occurs at α∗ = 1. Optimality here β 2 1 2 refers to minimizing the variance. Hint: Read the paper by Samuel-Cahn (1994).


Problems

69

 ,β  ) = var(β  ) > 0, and that ρ = (the correlation coefficient of (b) Similarly, show that cov(β 1 3 1 13 1 1     β 1 and β 3 ) = [var(β 1 )/var(β 3 )] 2 = (1 − ρ212 ) 2 with 0 < ρ13 < 1. Conclude that the optimal  is again α∗ = 1.  and β combination β 1 3    and β  is β  = (1 − ρ2 )β  + (c) Show that cov(β 2 , β 3 ) = 0 and that optimal combination of β 2 3 12 3 2   ρ12 β 2 = β 1 . This exercise demonstrates a more general result, namely that the BLUE of  , has a positive correlation with any other linear unbiased estimator of β, β in this case β 1 and that this correlation can be easily computed from the ratio of the variances of these two estimators.  denote the Best Linear Unbiased 12. Efficiency as Correlation. This is based on Oksanen (1993). Let β  denote any linear unbiased estimator of β. Show that the relative efficiency Estimator of β and let β  with respect to β  is the squared correlation coefficient between β  and β.  Hint: Compute the of β     is BLUE. This variance of β + λ(β − β) for any λ. This variance is minimized at λ = 0 since β 2 β)  which in turn proves the required result, see Zheng  ) = E(β should give you the result that E(β (1994).

13. For the numerical illustration given in section 3.9, what happens to the least squares regression 2   αOLS ) and se(β  OLS coefficient estimates ( αOLS , β OLS ), s , the estimated se( OLS ), t-statistic for α a b 2  and β for H ; α = 0, and H ; β = 0 and R when: OLS 0 0 (a) Yi is regressed on Xi + 5 rather than Xi . In other words, we add a constant 5 to each observation of the explanatory variable Xi and rerun the regression. It is very instructive to see how the computations in Table 3.1 are affected by this simple transformation on Xi .

(b) Yi + 2 is regressed on Xi . In other words, a constant 2 is added to Yi . (c) Yi is regressed on 2Xi . (A constant 2 is multiplied by Xi ). 14. For the cigarette consumption data given in Table 3.2. (a) Give the descriptive statistics for logC, logP and logY . Plot their histogram. Also, plot logC versus logY and logC versus logP . Obtain the correlation matrix of these variables. (b) Run the regression of logC on logY . What is the income elasticity estimate? What is its standard error? Test the null hypothesis that this elasticity is zero. What is the s and R2 of this regression? (c) Show that the square of the simple correlation coefficient between logC and logY is equal to R2 . Show that the square of the correlation coefficient between the fitted and actual values of logC is also equal to R2 . (d) Plot the residuals versus income. Also, plot the fitted values along with their 95% confidence band. 15. Consider the simple regression with no constant: Yi = βXi + ui

i = 1, 2, . . . , n

where ui ∼ IID(0, σ 2 ) independent of Xi . Theil (1971) showed that among all linear estimators in  − β)2 is given by Yi , the minimum mean square estimator for β, i.e., that which minimizes E(β  = β 2 n Xi Yi /(β 2 n X 2 + σ 2 ). β i i=1 i=1  = β/(1 + c), where c = σ 2 /β 2 n X 2 > 0. (a) Show that E(β) i=1 i   (b) Conclude that the Bias (β) = E(β) − β = −[c/(1 + c)]β. Note that this bias is positive  is biased towards zero. (negative) when β is negative (positive). This also means that β  n 2 2 2  = E(β  − β)2 = σ 2 /[ (c) Show that MSE(β) i=1 Xi + (σ /β )]. Conclude that it is smaller than  the MSE(β OLS ).


70

Chapter 3: Simple Linear Regression

Table 3.4 Energy Data for 20 countries RGDP (in 106 1975 U.S.$’s)

Country Malta Iceland Cyprus Ireland norway Finland Portugal Denmark Greece Switzerland Austria Sweden Belgium Netherlands Turkey Spain Italy U.K. France W. Germany

1251 1331 2003 11788 27914 28388 30642 34540 38039 42238 45451 59350 62049 82804 91946 159602 265863 279191 358675 428888

EN 106 Kilograms Coal Equivalents 456 1124 1211 11053 26086 26405 12080 27049 20119 23234 30633 45132 58894 84416 32619 88148 192453 268056 233907 352677

16. Table 3.4 gives cross-section Data for 1980 on real gross domestic product (RGDP) and aggregate energy consumption (EN ) for 20 countries (a) Enter the data and provide descriptive statistics. Plot the histograms for RGDP and EN. Plot EN versus RGDP. (b) Estimate the regression: log(En) = α + βlog(RGDP ) + u. Be sure to plot the residuals. What do they show? (c) Test H0 ; β = 1. (d) One of your Energy data observations has a misplaced decimal. Multiply it by 1000. Now repeat parts (a), (b) and (c). (e) Was there any reason for ordering the data from the lowest to highest energy consumption? Explain. Lesson Learned : Always plot the residuals. Always check your data very carefully. 17. Using the Energy Data given in Table 3.4, corrected as in problem 16 part (d), is it legitimate to reverse the form of the equation? log(RDGP ) = γ + δlog(En) + ǫ (a) Economically, does this change the interpretation of the equation? Explain. (b) Estimate this equation and compare R2 of this equation with that of the previous problem.  Why are they different? Also, check if  δ = 1/β.


References

71

(c) Statistically, by reversing the equation, which assumptions do we violate?  = R2 . (d) Show that  δβ (e) Effects of changing units in which variables are measured. Suppose you measured energy in BTU’s instead of kilograms of coal equivalents so that the original series was multiplied by 60. How does it change α and β in the following equations? log(En) = α + βlog(RDGP ) + u

En = α∗ + β ∗ RGDP + ν

 for the log-log model, whereas both α Can you explain why α  changed, but not β  ∗ and ∗  β changed for the linear model? (f) For the log-log specification and the linear specification, compare the GDP elasticity for Malta and W. Germany. Are both equally plausible? (g) Plot the residuals from both linear and log-log models. What do you observe? (h) Can you compare the R2 and standard errors from both models in part (g)? Hint: Retrieve  log(En) and log(En) in the log-log equation, exponentiate, then compute the residuals and s. These are comparable to those obtained from the linear model.

18. For the model considered in problem 16: log(En) = α + βlog(RGDP ) + u and measuring energy in BTU’s (like part (e) of problem 17). (a) What is the 95% confidence prediction interval at the sample mean? (b) What is the 95% confidence prediction interval for Malta? (c) What is the 95% confidence prediction interval for West Germany?

References Additional readings on the material covered in this chapter can be found in: Baltagi, B.H. (1995), “Optimal Weighting of Unbiased Estimators,” Econometric Theory, Problem 95.3.1, 11:637. Baltagi, B.H. and D. Levin (1992), “Cigarette Taxation: Raising Revenues and Reducing Consumption,” Structural Change and Economic Dynamics, 3: 321-335. Belsley, D.A., E. Kuh and R.E. Welsch (1980), Regression Diagnostics (Wiley: New York). Greene, W. (1993), Econometric Analysis (Macmillian: New York). Gujarati, D. (1995), Basic Econometrics (McGraw-Hill: New York). Johnston, J. (1984), Econometric Methods (McGraw-Hill: New York). Kelejian, H. and W. Oates (1989), Introduction to Econometrics (Harper and Row: New York). Kennedy, P. (1992), A Guide to Econometrics (MIT Press: Cambridge). Kmenta, J. (1986), Elements of Econometrics (Macmillan: New York). Maddala, G.S. (1992), Introduction to Econometrics (Macmillan: New York). Oksanen, E.H. (1993), “Efficiency as Correlation,” Econometric Theory, Problem 93.1.3, 9: 146. Samuel-Cahn, E. (1994), “Combining Unbiased Estimators,” The American Statistician, 48: 34-36. Wallace, D. and L. Silver (1988), Econometrics: An Introduction (Addison Wesley: New York). Zheng, J.X. (1994), “Efficiency as Correlation,” Econometric Theory, Solution 93.1.3, 10: 228.


72

Chapter 3: Simple Linear Regression

Appendix Centered and Uncentered R2 From the OLS regression on (3.1) we get Yi = Yi + ei

i = 1, 2, . . . , n

(A.1)

  OLS + Xi β where Yi = α OLS . Squaring and summing the above equation we get n n  2 n 2 2 i=1 Yi = i=1 Yi + i=1 ei

since

n



i=1 Yi ei

(A.2)

= 0. The uncentered R2 is given by

uncentered R2 = 1 −

n

2 n 2 i=1 ei / i=1 Yi

=

 2 2 i=1 Yi i=1 Yi /

n

(A.3)

Note that the total sum of squares for Yi is not expressed in deviation from the sample mean Y¯ . In other words, the uncentered R2 is the proportion of variation of ni=1 Yi2 that is explained by the regression on X.Regression usually report the centered R2 which was defined npackages n 2 2 in section 3.6 as 1 − ( i=1 ei / i=1 yi ) where yi = Yi − Y¯ . The latter measure focuses on explaining the variation in Yi after fitting the constant. From section 3.6, we have seen that a naive model with only a constant in it gives Y¯ as the estimate of the constant, see  also problem 2. The variation in Yi that is not explained by this naive model is ni=1 yi2 = ni=1 (Yi − Y¯ )2 . Subtracting nY¯ 2 from both sides of (A.2) we get n

2 i=1 yi

=

n

2 i=1 Yi

and the centered R2 is

 − nY¯ 2 + ni=1 e2i

    centered R2 = 1 − ( ni=1 e2i / ni=1 yi2 ) = ( ni=1 Yi2 − nY¯ 2 )/ ni=1 yi2

(A.4)

  If there is a constant in the model Y¯ = Y , see section 3.6, and ni=1 yi2 = ni=1 (Yi − Y )2 = n  2   n 2 ¯2 i2 / ni=1 yi2 which is the R2 reported by i=1 Yi − nY . Therefore, the centered R = i=1 y regression packages. If there is no constant in the model, some regression packages give you the option of (no constant) and the R2 reported is usually the uncentered R2 . Check your regression package documentation to verify what you are getting. We will encounter uncentered R2 again in constructing test statistics using regressions, see for example Chapter 11.


CHAPTER 4

Multiple Regression Analysis 4.1

Introduction

So far we have considered only one regressor X besides the constant in the regression equation. Economic relationships usually include more than one regressor. For example, a demand equation for a product will usually include real price of that product in addition to real income as well as real price of a competitive product and the advertising expenditures on this product. In this case Yi = α + β 2 X2i + β 3 X3i + .. + β K XKi + ui

i = 1, 2, . . . , n

(4.1)

where Yi denotes the i-th observation on the dependent variable Y, in this case the sales of this product. Xki denotes the i-th observation on the independent variable Xk for k = 2, . . . , K in this case, own price, the competitor’s price and advertising expenditures. α is the intercept and β 2 , β 3 , . . . , β K are the (K − 1) slope coefficients. The ui ’s satisfy the classical assumptions 1-4 given in Chapter 3. Assumption 4 is modified to include all the X’s appearing in the regression, i.e., every Xk for k = 2, . . . , K, is uncorrelated with the ui ’s with the property that n n ¯ 2 ¯ i=1 (Xki − Xk ) /n where Xk = i=1 Xki /n has a finite probability limit which is different from zero. Section 4.2 derives the OLS normal equations of this multiple regression model and discovers that an additional assumption is needed for these equations to yield a unique solution.

4.2

Least Squares Estimation

As explained in Chapter 3, least squares minimizes the residual sum of squares where the   Xki and α  denote guesses on the residuals are now given by ei = Yi − α − K β  and β k k=2 k regression parameters α and β k , respectively. The residual sum of squares RSS =

n

2 i=1 ei

=

n

i=1 (Yi

 X2i − .. − β  XKi )2 −α −β 2 K

is minimized by the following K first-order conditions:   ∂( ni=1 e2i )/∂ α = −2 ni=1 ei = 0   = −2 n ei Xki = 0, for k = 2, . . . , K. ∂( ni=1 e2i )/∂ β k i=1

or, equivalently n

K n Xki 2 n X2i + .. + β n + β i=1 i=1 i=1 Yi = α n n n 2    i=1 X2i + β 2 i=1 X2i + .. + β K i=1 Yi X2i = α

n

.. .

i=1 Yi XKi

=α 

n

.. .

i=1 XKi

.. .

 +β 2

n

i=1 X2i XKi

n

i=1 X2i XKi

.. .

 n X 2 + .. + β K i=1 Ki

(4.2)

(4.3)


74

Chapter 4: Multiple Regression Analysis

where the first equation multiplies the regression equation by the constant and sums, the second equation multiplies the regression equation  by X2 and sums,  and the K-th equation multiplies the regression equation by XK and sums. ni=1 ui = 0 and ni=1 ui Xki = 0 for k = 2, . . . , K are implicitly imposed to arrive at (4.3). Solving these K equations in K unknowns, we get the OLS estimators. This can be done more succinctly in matrix form, see Chapter 7. Assumptions 1-4 insure that the OLS estimator is BLUE. Assumption 5 introduces normality and as a result the OLS estimator is also (i) a maximum likelihood estimator, (ii) it is normally distributed, and (iii) it is minimum variance unbiased. Normality also allows test of hypotheses. Without the normality assumption, one has to appeal to the Central Limit Theorem and the fact that the sample is large to perform hypotheses testing. In order to make sure we can solve for the OLS estimators in (4.3) we need to impose one further assumption on the model besides those considered in Chapter 3. Assumption 6: No perfect multicollinearity, i.e., the explanatory variables are not perfectly correlated with each other. This assumption states that, no explanatory variable Xk for k = 2, . . . , K is a perfect linear combination of the other X’s. If assumption 6 is violated, then one of the equations in (4.2) or (4.3) becomes redundant and we would have K − 1 linearly independent equations in K unknowns. This means that we cannot solve uniquely for the OLS estimators of the K coefficients. Example 1: If X2i = 3X4i − 2X5i + X7i for i = 1, . . . , n, then multiplying this relationship by ei and summing over i we get n

i=1 X2i ei

=3

n

i=1 X4i ei

−2

n

i=1 X5i ei

+

n

i=1 X7i ei .

This means that the second OLS normal equation in (4.2) can be represented as a perfect linear combination of the fourth, fifth and seventh OLS normal equations. Knowing the latter three equations, the second equation adds no new information. Alternatively, one could substitute this relationship in the original regression equation (4.1). After some algebra, X2 would be eliminated and the resulting equation becomes: Yi = α + β 3 X3i + (3β 2 + β 4 )X4i + (β 5 − 2β 2 )X5i + β 6 X6i + (β 2 + β 7 )X7i

(4.4)

+.. + β K XKi + ui .

Note that the coefficients of X4i , X5i and X7i are now (3β 2 + β 4 ), (β 5 − 2β 2 ) and (β 2 + β 7 ), respectively. All of which are contaminated by β 2 . These linear combinations of β 2 , β 4 , β 5 and β 7 can be estimated from regression (4.4) which excludes X2i . In fact, the other X’s, not contaminated by this perfect linear relationship, will have coefficients that are not contaminated by β 2 and hence are themselves estimable using OLS. However, β 2 , β 4 , β 5 and β 7 cannot be estimated separately. Perfect multicollinearity means that we cannot separate the influence on Y of the independent variables that are perfectly related. Hence, assumption 6 of no perfect multicollinearity is needed to guarantee a unique solution of the OLS normal equations. Note that it applies to perfect linear relationships and does not apply to perfect non-linear relation2 like (years ships among the independent variables. In other words, one can include X1i and X1i of experience) and (years of experience)2 in an equation explaining earnings of individuals. Although, there is a perfect quadratic relationship between these independent variables, this is not a perfect linear relationship and therefore, does not cause perfect multicollinearity.


4.3

4.3

Residual Interpretation of Multiple Regression Estimates

75

Residual Interpretation of Multiple Regression Estimates

Although we did not derive an explicit solution for the OLS estimators of the β’s, we know  , the that they are the solutions to (4.2) or (4.3). Let us focus on one of these estimators, say β 2 OLS estimator of β 2 , the partial derivative of Yi with respect to X2i . As a solution to (4.2) or  is a multiple regression coefficient estimate of β . Alternatively, we can interpret β  (4.3), β 2 2 2 as a simple linear regression coefficient.

Claim 1: (i) Run the regression of X2 on all the other X’s in (4.1), and obtain the residuals 2 +  ν2 , i.e., X2 = X ν 2 . (ii) Run the simple regression of Y on  ν 2 , the resulting estimate of the  . slope coefficient is β 2 The first regression essentially cleans out the effect of the other X’s from X2 , leaving the vari can be interpreted as a simple linear regression ation unique to X2 in  ν 2 . Claim 1 states that β 2 coefficient of Y on this residual. This is in line with the partial derivative interpretation of β 2 . The proof of claim 1 is given in the Appendix. Using the results of the simple regression given ν 2 , we get in (3.4) with the regressor Xi replaced by the residual  n  = n  β ν 22i 2 i=1 ν 2 Yi / i=1 

(4.5)

and from (3.6) we get  ) = σ2/ var(β 2

n

ν 22i i=1 

(4.6)

 as a simple regression coefficient is the following: An alternative interpretation of β 2

Claim 2: (i) Run Y on all the other X’s and get the predicted Y and the residuals, say ω  . (ii)  is the resulting estimate of the slope coefficient. Run the simple linear regression of ω  on  ν 2. β 2 This regression cleans both Y and X2 from the effect of the other X’s and then regresses the cleaned out residuals of Y on those of X2 . Once again this is in line with the partial derivative interpretation of β 2 . The proof of claim 2 is simple and is given in the Appendix.  are important in that they provide an easy way of looking at These two interpretations of β 2 a multiple regression in the context of a simple linear regression. Also, it says that there is no need to clean the effects of one X from the other X’s to find its unique effect on Y . All one has to do is to include all these X’s in the same multiple regression. Problem 1 verifies this result with an empirical example. This will also be proved using matrix algebra in Chapter 7. Recall that R2 = 1 − RSS/T SS for any regression. Let R22 be the R2 for the regression n 2 n 2 ¯ 2 and of X2  on all the other X’s,then R2 = 1 − i=1 ν 2i / i=1 x22i wherex2i = X2i − X n n n n 2 2 2 ¯ ¯ X X2i /n; T SS = i=1 (X2i − X2 ) = i=1 x2i and RSS = i=1  ν 2i . Equivalently, 2n= 2 i=1  n 2 (1 − R2 ) and the = x ν  2i 2 2i i=1 i=1  ) = σ2/ var(β 2

n

ν 22i i=1 

= σ2 /

n

2 i=1 x2i (1

− R22 )

(4.7)

 ) holding σ 2 This means that the larger R22 , the smaller is (1 − R22 ) and the larger is var(β 2 n 2 and i=1 x2i fixed. This shows the relationship between multicollinearity and the variance of the OLS estimates. High multicollinearity between X2 and the other X’s will result in high  . Perfect multicollinearity is the extreme case R22 which in turn implies high variance for β 2 2  . In general, high multicollinearity where R2 = 1. This in turn implies an infinite variance for β 2 among the regressors yields imprecise estimates for these highly correlated variables. The least


76

Chapter 4: Multiple Regression Analysis

squares regression estimates are still unbiased as long as assumptions 1 and 4 are satisfied, but these estimates are unreliable asreflected by their high variances. However, it is important n x2 could counteract the effect of a high R2 leading to note that a low σ 2 and a high 2 i=1 2i  . Maddala (2001) argues that high intercorrelation among the to a significant t-statistic for β 2 explanatory variables are neither necessary nor sufficient to cause the multicollinearity problem. In practice, multicollinearity is sensitive to the addition or deletion of observations. More on this in Chapter 8. Looking at high intercorrelations among the explanatory variables is useful only as a complaint. It is more important to look at the standard errors and t-statistics to assess the seriousness of multicollinearity. Much has been written on possible solutions to the multicollinearity problem, see Hill and Adkins (2001) for a recent summary. Credible candidates include: (i) obtaining new and better data, but this is rarely available; (ii) introducing nonsample information about the model parameters based on previous empirical research or economic theory. The problem with the latter solution is that we never truly know whether the information we introduce is good enough to reduce estimator Mean Square Error.

4.4

Overspecification and Underspecification of the Regression Equation

So far we have assumed that the true linear regression relationship is always correctly specified. This is likely to be violated in practice. In order to keep things simple, we consider the case where the true model is a simple regression with one regressor X1 . True model: Yi = α + β 1 X1i + ui with ui ∼ IID(0, σ 2 ), but the estimated model is overspecified with the inclusion of an additional irrelevant variable X2 , i.e.,  X1i + β  X2i Estimated model: Yi = α +β 1 2

n  = n  ν 21i where  ν 1 is the OLS From the previous section, it is clear that β 1 i=1  i=1 ν 1i Yi / residuals of X1 on X2 . Substituting the true model for Y we get n  = β n ν1i X1i / n ν2 + n  ν 21i β 1 1 i=1  i=1 ν 1i ui / i=1 1i i=1

n

  1i +  1i  = 0. But, X1i = X ν 1i and ni=1 X ν 1i = 0 implying that ni=1  ν 1i X1i = 2 . Hence, ν  i=1 1i

since  n

ν 1i i=1 

n  = β + n  β ν 21i 1 1 i=1 ν 1i ui / i=1 

(4.8)

 ) = β since  and E(β ν 1 is a linear combination of the X’s, and E(Xk u) = 0 for k = 1, 2. Also, 1 1  ) = σ2/ var(β 1

n

ν 21i i=1 

= σ2 /

n

2 i=1 x1i (1

− R12 )

(4.9)

¯ 1 and R2 is the R2 of the regression of X1 on X2 . Using the true model where x1i = X1i − X 1 n n 2 to estimate β 1 , one would get b1 = i=1 x1i yi / i=1 x1i with E(b1 ) = β 1 and var(b1 ) =


4.4

Overspecification and Underspecification of the Regression Equation

77

  1 ) ≥ var(b1 ). Note also that in the overspecified model, the estimate σ 2 / ni=1 x21i . Hence, var(β for β 2 which has a true value of zero is given by n  = n  β ν 22i 2 i=1 ν 2i Yi / i=1 

(4.10)

where  ν 2 is the OLS residual of X2 on X1 . Substituting the true model for Y we get n  = n  β ν 22i 2 i=1 ν 2i ui / i=1 

(4.11)

   ) = 0 since  ν 2i X1i = 0 and ni=1 ν2i = 0. Hence, E(β ν 2 is a linear combination since ni=1  2 of the X’s and E(Xk u) = 0 for k = 1, 2. In summary, overspecification still yields unbiased estimates of β 1 and β 2 , but the price is a higher variance. Similarly, the true model could be a two-regressors model True model: Yi = α + β 1 X1i + β 2 X2i + ui where ui ∼ IID(0, σ 2 ) but the estimated model is  X1i Estimated model: Yi = α +β 1

The estimated model omits a relevant variable X2 and underspecifies the true relationship. In this case  = n x1i Yi / n x2 β 1 i=1 i=1 1i

(4.12)

¯ 1 . Substituting the true model for Y we get where x1i = X1i − X 1 = β 1 + β 2 n x1i X2i / n x2 + n x1i ui / n x2 β i=1 i=1 1i i=1 i=1 1i

(4.13)

 ) = β + β b12 since E(x1 u) = 0 with b12 = n x1i X2i / n x2 . Note that b12 Hence, E(β 1 1 2 i=1 i=1 1i is the regression slope estimate obtained by regressing X2 on X1 and a constant. Also, the  ) = E(β  − E(β  ))2 = E(n x1i ui / n x2 )2 = σ 2 / n x2 var(β 1 1 1 i=1 i=1 1i i=1 1i

which understates the variance of the estimate of β 1 obtained from the true model, i.e., b1 =   n 1i Yi / ni=1  ν 21i with i=1 ν var(b1 ) = σ 2 /

n

ν 21i i=1 

= σ2/

n

2 i=1 x1i (1

 ). − R12 ) ≥ var(β 1

(4.14)

In summary, underspecification yields biased estimates of the regression coefficients and understates the variance of these estimates. This is also an example of imposing a zero restriction on β 2 when in fact it is not true. This introduces bias, because the restriction is wrong, but reduces the variance because it imposes more information even if this information may be false. We will encounter this general principle again when we discuss distributed lags in Chapter 6.


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Chapter 4: Multiple Regression Analysis

4.5

R-Squared versus R-Bar-Squared

Since OLS minimizes the residual sums of squares, adding one or more variables to the regression cannot increase this residual sums of squares. After all, we are minimizing over a larger dimension parameter set and the minimum there is smaller or equal to that over a subset of the parameter space, , adding more  vari see problem 4. Therefore, for the same dependent variable Y  ables makes ni=1 e2i non-increasing and R2 non-decreasing, since R2 = 1 − ( ni=1 e2i / ni=1 yi2 ). Hence, a criteria of selecting a regression that “maximizes R2 ” does not make sense, since we can add more variables to this regression and improve on this R2 (or at worst leave it the same). In order to penalize the researcher for adding an extra variable, one computes ¯ 2 = 1 − [n e2i /(n − K)]/[n yi2 /(n − 1)] R (4.15) i=1 i=1 n n 2 2 where i=1 yi have been adjusted by their i=1 ei and  degrees of freedom. Note that the 2 numerator is the s of the regression and is equal to ni=1 e2i /(n − K). This differs from the s2 in Chapter 3 in the degrees of freedom. Here, it is n − K, because we have estimated K coefficients, or because (4.2) represents K relationships among the residuals. Therefore knowing  (n − K) residuals we can deduce the other K residuals from (4.2). ni=1 e2i is non-increasing as we add more variables, but the degrees of freedom decrease by one with every added variable. Therefore, s2 will decrease only if the effect of the ni=1 e2i decrease outweighs the effect of the ¯ 2 , i.e., penalizing each added one degree of freedom loss on s2 . This is exactly the idea behind R ¯2 variable by decreasing n the2degrees of freedom by one. Hence,2 this variable will increase R only if the reduction in i=1 ei outweighs this loss, i.e., only if s is decreased. Using the definition ¯ 2 , one can relate it to R2 as follows: of R ¯ 2 ) = (1 − R2 )[(n − 1)/(n − K)] (1 − R

4.6

(4.16)

Testing Linear Restrictions

In the simple linear regression chapter, we proved that the OLS estimates are BLUE provided assumptions 1 to 4 were satisfied. Then we imposed normality on the disturbances, assumption 5, and proved that the OLS estimators are in fact the maximum likelihood estimators. Then we derived the Cram´er-Rao lower bound, and proved that these estimates are efficient. This will be done in matrix form in Chapter 7 for the multiple regression case. Under normality one can test hypotheses about the regression. Basically, any regression package will report the OLS estimates, their standard errors and the corresponding t-statistics for the null hypothesis that each individual coefficient is zero. These are tests of significance for each coefficient separately. But one may be interested in a joint test of significance for two or more coefficients simultaneously, or simply testing whether linear restrictions on the coefficients of the regression are satisfied. This will be developed more formally in Chapter 7. For now, all we assume is that the reader can perform regressions using his or her favorite software like EViews, STATA, SAS, TSP, SHAZAM, LIMDEP or GAUSS. The solutions to (4.2) or (4.3) result in the OLS estimates. These multiple regression coefficient estimates can be interpreted as simple regression estimates as shown in section 4.3. This allows a simple derivation of their standard errors. Now, we would like to use these regressions to test linear restrictions. The strategy followed is to impose these restrictions on the model and run the resulting restricted regression. The


4.6

Testing Linear Restrictions

79

corresponding Restricted Residual Sums of Squares is denoted by RRSS. Next, one runs the regression without imposing these linear restrictions to obtain the Unrestricted Residual Sums of Squares, which we denote by URSS. Finally, one forms the following F -statistic: F =

(RRSS − URSS )/ℓ ∼ Fℓ,n−K URSS /(n − K)

(4.17)

where ℓ denotes the number of restrictions, and n − K gives the degrees of freedom of the unrestricted model. The idea behind this test is intuitive. If the restrictions are true, then the RRSS should not be much different from the URSS. If RRSS is different from URSS, then we reject these restrictions. The denominator of the F -statistic is a consistent estimate of the unrestricted regression variance. Dividing by the latter makes the F -statistic invariant to units of measurement. Let us consider two examples: Example 2: Testing the joint significance of two regression coefficients. For e.g., let us test the following null hypothesis H0 ; β 2 = β 3 = 0. These are two restrictions β 2 = 0 and β 3 = 0 and they are to be tested jointly. We know how to test for β 2 = 0 alone or β 3 = 0 alone with individual t-tests. This is a test of joint significance of the two coefficients. Imposing this restriction, means the removal of X2 and X3 from the regression, i.e., running the regression of Y on X4 , . . . , XK excluding X2 and X3 . Hence, the number of parameters to be estimated becomes (K − 2) and the degrees of freedom of this restricted regression are n − (K − 2). The unrestricted regression is the one including all the X’s in the model. Its degrees of freedom are (n − K). The number of restrictions are 2 and this can also be inferred from the difference between the degrees of freedom of the restricted and unrestricted regressions. All the ingredients are now available for computing F in (4.17) and this will be distributed as F2,n−K . Example 3: Test the equality of two regression coefficients H0 ; β 3 = β 4 against the alternative that H1 ; β 3 = β 4 . Note that H0 can be rewritten as H0 ; β 3 − β 4 = 0. This can be tested using a t-statistic that tests whether d = β 3 − β 4 is equal to zero. From the unrestricted regression, we  −β  with var(d)  = var(β  )+var(β  ) − 2cov(β  ,β  ). The variance-covariance can obtain d = β 3 4 3 4 3 4 matrix of the regression coefficients can be printed out with any regression package. In section 4.3, we gave these variances and covariances a simple regression interpretation. This means   and the t-statistic is simply t = (d − 0)/se(d)  which is distributed as  that se(d) = var(d)

tn−K under H0 . Alternatively, one can run an F -test with the RRSS obtained from running the following regression Yi = α + β 2 X2i + β 3i (X3i + X4i ) + β 5 X5i + .. + β K XKi + ui with β 3 = β 4 substituted in for β 4 . This regression has the variable (X3i + X4i ) rather than X3i and X4i separately. The URSS is the regression of Y on all the X’s in the model. The degrees of freedom of the resulting F -statistic are 1 and n − K. The numerator degree of freedom states that there is only one restriction. It will be proved in Chapter 7 that the square of the t-statistic is exactly equal to the F -statistic just derived. Both methods of testing are equivalent. The first one computes only the unrestricted regression and involves some further variance computations, while the latter involves running two regressions and computing the usual F -statistic. Example 4: Test the joint hypothesis H0 ; β 3 = 1 and β 2 − 2β 4 = 0. These two restrictions are usually obtained from prior information or imposed by theory. The first restriction is β 3 = 1.


80

Chapter 4: Multiple Regression Analysis

The value 1 could have been any other constant. The second restriction shows that a linear combination of β 2 and β 4 is equal to zero. Substituting these restrictions in (4.1) we get Yi = α + β 2 X2i + X3i + 12 β 2 X4i + β 5 X5i + .. + β K XKi + ui which can be written as Yi − X3i = α + β 2 (X2i + 12 X4i ) + β 5 X5i + .. + β K XKi + ui Therefore, the RRSS can be obtained by regressing (Y − X3 ) on (X2 + 21 X4 ), X5 , . . . , XK . This regression has n − (K − 2) degrees of freedom. The URSS is the regression with all the X’s included. The resulting F -statistic has 2 and n − K degrees of freedom. Example 5: Testing constant returns to scale in a Cobb-Douglas production function. Q = AK α Lβ E γ M δ eu is a Cobb-Douglas production function with capital(K), labor(L), energy(E) and material(M ). Constant returns to scale means that a proportional increase in the inputs produces the same proportional increase in output. Let this proportional increase be λ, then K ∗ = λK, L∗ = λL, E ∗ = λE and M ∗ = λM . Q∗ = λ(α+β+γ+δ) AK α Lβ E γ M δ eu = λ(α+β+γ+δ) Q. For this last term to be equal to λQ, the following restriction must hold: α + β + γ + δ = 1. Hence, a test of constant returns to scale is equivalent to testing H0 ; α + β + γ + δ = 1. The Cobb-Douglas production function is nonlinear in the variables, and can be linearized by taking logs of both sides, i.e., logQ = logA + αlogK + βlogL + γlogE + δlogM + u

(4.18)

This is a linear regression with Y = logQ, X2 = logK, X3 = logL, X4 = logE and X5 = logM . Ordinary least squares is BLUE on this non-linear model as long as u satisfies assumptions 1-4. Note that these disturbances entered the original Cobb-Douglas production function multiplicatively as exp(ui ). Had these disturbances entered additively as Q = AK α Lβ E γ M δ + u then taking logs does not simplify the right hand side and one has to estimate this with nonlinear least squares, see Chapter 8. Now we can test constant returns to scale as follows. The unrestricted regression is given by (4.18) and its degrees of freedom are n − 5. Imposing H0 means substituting the linear restriction by replacing say β by (1 − α − γ − δ). This results after collecting terms in the following restricted regression with one less parameter log(Q/L) = logA + αlog(K/L) + γlog(E/L) + δlog(M/L) + u

(4.19)

The degrees of freedom are n − 4. Once again all the ingredients for the test in (4.17) are there and this statistic is distributed as F1 , n − 5 under the null hypothesis. Example 6: Joint significance of all the slope coefficients. The null hypothesis is H0 ; β 2 = β 3 = .. = β K = 0 against the alternative H1 ; at least one β k = 0 for k = 2, . . . , K. Under the null, only the constant is left in the regression. Problem 3.2 showed that for a regression of Y on a constant only, the least estimate of α is Y¯ . This means that the corresponding residual sum of n squares 2 ¯ squares is i=1 (Yi −Y ) . Therefore, RRSS = Total sums of squares of regression (4.1) = Σni=1 yi2 .


4.7

Dummy Variables

81

 The URSS is the usual residual sums of squares ni=1 e2i from the unrestricted regression given by (4.1). Hence, the corresponding F -statistic for H0 is   ( ni=1 yi2 − ni=1 e2i )/(K − 1) R2 n−K (T SS − RSS)/(K − 1) n 2 = = · (4.20) F = 2 RSS/(n − K) 1−R K −1 i=1 ei /(n − K)   where R2 = 1−( ni=1 e2i / ni=1 yi2 ). This F -statistic has (K −1) and (n−K) degrees of freedom under H0 , and is usually reported by regression packages.

4.7

Dummy Variables

Many explanatory variables are qualitative in nature. For example, the head of a household could be male or female, white or non-white, employed or unemployed. In this case, one codes these variables as “M ” for male and “F ” for female, or change this qualitative variable into a quantitative variable called FEMALE which takes the value “0” for male and “1” for female. This obviously begs the question: “why not have a variable MALE that takes on the value 1 for male and 0 for female?” Actually, the variable MALE would be exactly 1-FEMALE. In other words, the zero and one can be thought of as a switch, which turns on when it is 1 and off when it is 0. Suppose that we are interested in the earnings of households, denoted by EARN, and MALE and FEMALE are the only explanatory variables available, then problem 10 asks the reader to verify that running OLS on the following model: EARN = αM MALE + αF FEMALE + u

(4.21)

gives α  M = “average earnings of the males in the sample” and α  F = “average earnings of the females in the sample.” Notice that there is no intercept in (4.21), this is because of what is known in the literature as the “dummy variable trap.” Briefly stated, there will be perfect multicollinearity between MALE, FEMALE and the constant. In fact, MALE + FEMALE = 1. Some researchers may choose to include the intercept and exclude one of the sex dummy variables, say MALE, then EARN = α + βFEMALE + u

(4.22)

= and the OLS estimates give α  = “average earnings of males in the sample” = α  M , while β α F − α  M = “the difference in average earnings between females and males in the sample.” Regression (4.22) is more popular when one is interested in contrasting the earnings between males and females and obtaining with one regression the markup or markdown in average earnings ( αF − α  M ) as well as the test of whether this difference is statistically different from zero.  in (4.22). On the other hand, if one is interested in This would be simply the t-statistic on β estimating the average earnings of males and females separately, then model (4.21) should be  M = 0 would involve further calculathe one to consider. In this case, the t-test for α F − α tions not directly given from the regression in (4.21) but similar to the calculations given in Example 3. What happens when another qualitative variable is included, to depict another classification of the individuals in the sample, say for example, race? If there are three race groups in the sample, WHITE, BLACK and HISPANIC. One could create a dummy variable for each of these classifications. For example, WHITE will take the value 1 when the individual is white


82

Chapter 4: Multiple Regression Analysis

and 0 when the individual is non-white. Note that the dummy variable trap does not allow the inclusion of all three categories as they sum up to 1. Also, even if the intercept is dropped, once MALE and FEMALE are included, perfect multicollinearity is still present because MALE + FEMALE = WHITE + BLACK + HISPANIC. Therefore, one category from race should be dropped. Suits (1984) argues that the researcher should use the dummy variable category omission to his or her advantage, in interpreting the results, keeping in mind the purpose of the study. For example, if one is interested in comparing earnings across the sexes holding race constant, the omission of MALE or FEMALE is natural, whereas, if one is interested in the race differential in earnings holding gender constant, one of the race variables should be omitted. Whichever variable is omitted, this becomes the base category for which the other earnings are compared. Most researchers prefer to keep an intercept, although regression packages allow for a no intercept option. In this case one should omit one category from each of the race and sex classifications. For example, if MALE and WHITE are omitted: EARN = α + β F FEMALE + β B BLACK + β H HISPANIC + u

(4.23)

Assuming the error u satisfies all the classical assumptions, and taking expected values of both sides of (4.23), one can see that the intercept α = the expected value of earnings of the omitted category which is “white males”. For this category, all the other switches are off. Similarly, α + β F is the expected value of earnings of “white females,” since the FEMALE switch is on. One can conclude that β F = difference in the expected value of earnings between white females and white males. Similarly, one can show that α + β B is the expected earnings of “black males” and α + β F + β B is the expected earnings of “black females.” Therefore, β F represents the difference in expected earnings between black females and black males. In fact, problem 11 asks the reader to show that β F represents the difference in expected earnings between hispanic females and hispanic males. In other words, β F represents the differential in expected earnings between females and males holding race constant. Similarly, one can show that β B is the difference in expected earnings between blacks and whites holding sex constant, and β H is the differential in expected earnings between hispanics and whites holding sex constant. The main key to the interpretation of the dummy variable coefficients is to be able to turn on and turn off the proper switches, and write the correct expectations. The real regression will contain other quantitative and qualitative variables, like EARN = α + β F FEMALE + β B BLACK + β H HISPANIC + γ 1 EXP

(4.24)

+γ 2 EXP 2 + γ 3 EDUC + γ 4 UNION + u where EXP is years of job experience, EDUC is years of education, and UNION is 1 if the individual belongs to a union and 0 otherwise. EXP2 is the squared value of EXP. Once again, one can interpret the coefficients of these regressions by turning on or off the proper switches. For example, γ 4 is interpreted as the expected difference in earnings between union and non-union members holding all other variables included in (4.24) constant. Halvorsen and Palmquist (1980) warn economists about the interpretation of dummy variable coefficients when the dependent variable is in logs. For example, if the earnings equation is semi-logarithmic: log(Earnings) = α + βUNION + γEDUC + u then γ = % change in earnings for one extra year of education, holding union membership constant. But, what about the returns for union membership? If we let Y1 = log(Earnings)


4.7

Dummy Variables

83

when the individual belongs to a union, and Y0 = log(Earnings) when the individual does not belong to a union, then g = % change in earnings due to union membership = (eY1 − eY0 )/eY0 . Equivalently, one can write that log(1 + g) = Y1 − Y0 = β, or that g = eβ − 1. In other words, one should not hasten to conclude that β has the same interpretation as γ. In fact, the %  change in earnings due to union membership is eβ − 1 and not β. The error involved in using β  β  rather than e − 1 to estimate g could be substantial, especially if β is large. For example, when  = 0.5, 0.75, 1; g = eβ − 1 = 0.65, 1.12, 1.72, respectively. Kennedy (1981) notes that if β  is β  implies consistency unbiased for β,  g is not necessarily unbiased for g. However, consistency of β   β+0.5Var( β) β . Based on this for g. If one assumes log-normal distributed errors, then E(e ) = e     is a consistent  β) result, Kennedy (1981) suggests estimating g by g = eβ+0.5Var(β) -1, where Var(  estimate of Var(β).

Another use of dummy variables is in taking into account seasonal factors, i.e., including 3 seasonal dummy variables with the omitted season becoming the base for comparison.1 For example: Sales = α + β W W inter + β S Spring + β F F all + γ 1 P rice + u

(4.25)

the omitted season being the Summer season, and if (4.25) models the sales of air-conditioning units, then β F is the difference in expected sales between the Fall and Summer seasons, holding the price of an air-conditioning unit constant. If these were heating units one may want to change the base season for comparison. Another use of dummy variables is for War years, where consumption is not at its normal level say due to rationing. Consider estimating the following consumption function Ct = α + βYt + δWARt + ut

t = 1, 2, . . . , T

(4.26)

where Ct denotes real per capita consumption, Yt denotes real per capita personal disposable income, and W ARt is a dummy variable taking the value 1 if it is a War time period and 0 otherwise. Note that the War years do not affect the slope of the consumption line with respect to income, only the intercept. The intercept is α in non-War years and α + δ in War years. In other words, the marginal propensity out of income is the same in War and non-War years, only the level of consumption is different. Of course, one can dummy other unusual years like periods of strike, years of natural disaster, earthquakes, floods, hurricanes, or external shocks beyond control, like the oil embargo of 1973. If this dummy includes only one year like 1973, then the dummy variable for 1973, call it D73 , takes the value 1 for 1973 and zero otherwise. Including D73 as an extra variable in the regression has the effect of removing the 1973 observation from estimation purposes, and the resulting regression coefficients estimates are exactly the same as those obtained excluding the 1973 observation and its corresponding dummy variable. In fact, using matrix algebra in Chapter 7, we will show that the coefficient estimate of D73 is the forecast error for 1973, using the regression that ignores the 1973 observations. In addition, the standard error of the dummy coefficient estimates is the standard error of this forecast. This is a much easier way of obtaining the forecast error and its standard error from the regression package without additional computations, see Salkever (1976). More on this in Chapter 7.


84

Chapter 4: Multiple Regression Analysis

Interaction Effects So far the dummy variables have been used to shift the intercept of the regression keeping the slopes constant. One can also use the dummy variables to shift the slopes by letting them interact with the explanatory variables. For example, consider the following earnings equation: EARN = α + αF FEMALE + βEDUC + u

(4.27)

In this regression, only the intercept shifts from males to females. The returns to an extra year of education is simply β, which is assumed to be the same for males as well as females. But if we now introduce the interaction variable (FEMALE × EDUC), then the regression becomes: EARN = α + αF FEMALE + βEDUC + γ(FEMALE × EDUC ) + u

(4.28)

In this case, the returns to an extra year of education depends upon the sex of the individual. In fact, ∂(EARN )/∂(EDUC ) = β + γ(FEMALE ) = β if male, and β + γ if female. Note that the interaction variable = EDUC if the individual is female and 0 if the individual is male. Estimating (4.28) is equivalent to estimating two earnings equations, one for males and another one for females, separately. The only difference is that (4.28) imposes the same variance across the two groups, whereas separate regressions do not impose this, albeit restrictive, equality of the variances assumption. This set-up is ideal for testing the equality of slopes, equality of intercepts, or equality of both intercepts and slopes across the sexes. This can be done with the F -test described in (4.17). In fact, for H0 ; equality of slopes, given different intercepts, the restricted residuals sum of squares (RRSS) is obtained from (4.27), while the unrestricted residuals sum of squares (URSS) is obtained from (4.28). Problem 12 asks the reader to set up the F -test for the following null hypothesis: (i) equality of slopes and intercepts, and (ii) equality of intercepts given the same slopes. Dummy variables have many useful applications in economics. For example, several tests including the Chow (1960) test, and Utts (1982) Rainbow test described in Chapter 8, can be applied using dummy variable regressions. Additionally, they can be used in modeling splines, see Poirier (1976) and Suits, Mason and Chan (1978), and fixed effects in panel data, see Chapter 12. Finally, when the dependent variable is itself a dummy variable, the regression equation needs special treatment, see Chapter 13 on qualitative limited dependent variables. Empirical Example: Table 4.1 gives the results of a regression on 595 individuals drawn from the Panel Study of Income Dynamics (PSID) in 1982. This data is provided on the Springer web site as EARN.ASC. A description of the data is given in Cornwell and Rupert (1988). In particular, log wage is regressed on years of education (ED), weeks worked (WKS), years of full-time work experience (EXP), occupation (OCC = 1, if the individual is in a blue-collar occupation), residence (SOUTH = 1, SMSA = 1, if the individual resides in the South, or in a standard metropolitan statistical area), industry (IND = 1, if the individual works in a manufacturing industry), marital status (MS = 1, if the individual is married), sex and race (FEM = 1, BLK = 1, if the individual is female or black), union coverage (UNION = 1, if the individual’s wage is set by a union contract). These results show that the returns to an extra year of schooling is 5.7%, holding everything else constant. It shows that Males on the average earn more than Females. Blacks on the average earn less than Whites, and Union workers earn more than non-union workers. Individuals residing in the South earn less than those living elsewhere. Those residing in a standard metropolitan statistical area earn more on the average than those


Note

85

Table 4.1 Earnings Regression for 1982 Dependent Variable: LWAGE Analysis of Variance Source Model Error C Total

DF 12 582 594

Root MSE Dep Mean C.V.

Sum of Squares 52.48064 61.68465 114.16529

Mean Square 4.37339 0.10599

F Value 41.263

0.32556 6.95074 4.68377

R-square Adj R-sq

0.4597 0.4485

Prob > F 0.0001

Parameter Estimates

Variable INTERCEP WKS SOUTH SMSA MS EXP EXP2 OCC IND UNION FEM BLK ED

DF 1 1 1 1 1 1 1 1 1 1 1 1 1

Parameter

Standard

T for H0:

Estimate 5.590093 0.003413 –0.058763 0.166191 0.095237 0.029380 –0.000486 –0.161522 0.084663 0.106278 –0.324557 –0.190422 0.057194

Error 0.19011263 0.00267762 0.03090689 0.02955099 0.04892770 0.00652410 0.00012680 0.03690729 0.02916370 0.03167547 0.06072947 0.05441180 0.00659101

Parameter=0 29.404 1.275 –1.901 5.624 1.946 4.503 –3.833 –4.376 2.903 3.355 –5.344 –3.500 8.678

Prob > |T| 0.0001 0.2030 0.0578 0.0001 0.0521 0.0001 0.0001 0.0001 0.0038 0.0008 0.0001 0.0005 0.0001

who do not. Individuals who work in a manufacturing industry or are not blue collar workers or are married earn more on the average than those who are not. For EXP 2 = (EXP )2 , this regression indicates a significant quadratic relationship between earnings and experience. All the variables were significant at the 5% level except for WKS, SOUTH and MS.

Note 1. There are more sophisticated ways of seasonal adjustment than introducing seasonal dummies, see Judge et al. (1985).

Problems 1. For the Cigarette Data given in Table 3.2. Run the following regressions: (a) Real per capita consumption of cigarettes on real price and real per capita income. (All variables are in log form, and all regressions in this problem include a constant).


86

Chapter 4: Multiple Regression Analysis (b) Real per capita consumption of cigarettes on real price. (c) Real per capita income on real price. (d) Real per capita consumption on the residuals of part (c). (e) Residuals from part (b) on the residuals in part (c). (f) Compare the regression slope estimates in parts (d) and (e) with the regression coefficient estimate of the real income coefficient in part (a), what do you conclude? 2. Simple versus Multiple Regression Coefficients. This is based on Baltagi (1987b). Consider the multiple regression Yi = α + β 2 X2i + β 3 X3i + ui

i = 1, 2, . . . , n

along with the following auxiliary regressions: X2i X3i

=  a + bX3i +  ν 2i  2i +  =  c + dX ν 3i

 , the OLS estimate of β can be interpreted as a simple regression In section 4.3, we showed that β 2 2  . Kennedy (1981, p. 416) of Y on the OLS residuals  ν 2 . A similar interpretation can be given to β 3   claims that β 2 is not necessarily the same as δ 2 , the OLS estimate of δ 2 obtained from the regression Y on  ν2,  ν 3 and a constant, Yi = γ + δ 2  ν 2i + δ 3  ν 3i + wi . Prove this claim by finding a relationship  and the  between the β’s δ’s.

3. For the simple regression Yi = α + βXi + ui considered in Chapter 3, show that n n 2  (a) β OLS = i=1 xi can be obtained using the residual interpretation by regressing i=1 xi yi / X on a constant first, getting the residuals  ν and then regressing Y on  ν.  ¯ X can be obtained using the residual interpretation by regressing 1 on X (b) α  OLS = Y¯ − β OLS and obtaining the residuals ω  and then regressing Y on ω .  (c) Check the var( αOLS ) and var(β OLS ) in parts (a) and (b) with those obtained from the residualing interpretation. 4. Effect of Additional Regressors on R2 . This is based on Nieswiadomy (1986).

(a) Suppose that the multiple regression given in (4.1) has K1 regressors in it. Denote the least squares sum of squared errors by SSE1 . Now add K2 regressors so that the total number of regressors is K = K1 + K2 . Denote the corresponding least squares sum of squared errors by SSE2 . Show that SSE2 ≤ SSE1 , and conclude that the corresponding R-squares satisfy R22 ≥ R12 . ¯2. (b) Derive the equality given in (4.16) starting from the definition of R2 and R ¯ 2 when the F -statistic for the joint ¯ ¯2 ≥ R (c) Show that the corresponding R-squares satisfy R 2 1 significance of these additional K2 regressors is less than or equal to one. 5. Let Y be the output and X2 = skilled labor and X3 = unskilled labor in the following relationship: 2 2 + β 6 X3i + ui Yi = α + β 2 X2i + β 3 X3i + β 4 (X2i + X3i ) + β 5 X2i

What parameters are estimable by OLS? 6. Suppose that we have estimated the parameters of the multiple regression model: Yt = β 1 + β 2 Xt2 + β 3 Xt3 + ut by Ordinary Least Squares (OLS) method. Denote the estimated residuals by (et , t = 1, . . . , T ) and the predicted values by (Yt , t = 1, . . . , T ).


Problems

87

(a) What is the R2 of the regression of e on a constant, X2 and X3 ? (b) If we regress Y on a constant and Y , what are the estimated intercept and slope coefficients? What is the relationship between the R2 of this regression and the R2 of the original regression? (c) If we regress Y on a constant and e, what are the estimated intercept and slope coefficients? What is the relationship between the R2 of this regression and the R2 of the original regression? (d) Suppose that we add a new explanatory variable X4 to the original model and re-estimate the parameters by OLS. Show that the estimated coefficient of X4 and its estimated standard error will be the same as in the OLS regression of e on a constant, X2 , X3 and X4 . 7. Consider the Cobb-Douglas production function in example 5. How can you test for constant returns to scale using a t-statistic from the unrestricted regression given in (4.18). 8. For the multiple regression given in (4.1). Set up the F -statistic described in (4.17) for testing (a) H0 ; β 2 = β 4 = β 6 . (b) H0 ; β 2 = −β 3 and β 5 − β 6 = 1. 9. Monte Carlo Experiments. Hanushek and Jackson (1977, pp. 60-65) generated the following data Yi = 15 + 1X2i + 2X3i + ui for i = 1, 2, . . . , 25 with a fixed set of X2i and X3i , and ui ’s that are IID ∼ N (0, 100). For each set of 25 ui ’s drawn randomly from the normal distribution, a corresponding set of 25 Yi ’s are created from the above equation. Then OLS is performed on the resulting data set. This can be repeated as many times as we can afford. 400 replications were performed by Hanushek and Jackson. This means that they generated 400 data sets each of size 25 and ran 400 regressions giving 400 OLS estimates of α, β 2 , β 3 and σ 2 . The classical assumptions are satisfied for this model, by construction, so we expect these OLS estimators to be BLUE, MLE and efficient. (a) Replicate the Monte Carlo experiments of Hanushek and Jackson (1977) and generate the means of the 400 estimates of the regression coefficients as well as σ 2 . Are these estimates unbiased? (b) Compute the standard deviation of these 400 estimates and call this σ b . Also compute the average of the 400 standard errors of the regression estimates reported by the regression. Denote this mean by sb . Compare these two estimates of the standard deviation of the regression coefficient estimates to the true standard deviation knowing the true σ 2 . What do you conclude? (c) Plot the frequency of these regression coefficients estimates? Does it resemble its theoretical distribution. (d) Increase the sample size form 25 to 50 and repeat the experiment. What do you observe? 10.

(a) Derive the OLS estimates of αF and αM for Yi = αF Fi + αM Mi + ui where Y is Earnings, F is FEMALE and M is MALE, see (4.21). Show that α  F = Y¯F , the average of the Yi ’s only ¯ for females, and α  M = YM , the average of the Yi ’s only for males. (b) Suppose that the regression is Yi = α + βFi + ui , see (4.22). Show that α  = α  M , and =α β F − α M . (c) Substitute M = 1 − F in (4.21) and show that α = αM and β = αF − αM . (d) Verify parts (a), (b) and (c) using the earnings data underlying Table 4.1.

11. For equation (4.23) EARN = α + β F F EM ALE + β B BLACK + β H HISP AN IC + u Show that


88

Chapter 4: Multiple Regression Analysis (a) E(Earnings/Hispanic Female) = α + β F + β H ; also E(Earnings/Hispanic Male) = α + β H . Conclude that β F = E(Earnings/Hispanic Female) – E(Earnings/Hispanic Male). (b) E(Earnings/Hispanic Female) – E(Earnings/White Female) = E(Earnings/Hispanic Male) – E(Earnings/White Male) = β H . (c) E(Earnings/Black Female) – E(Earnings/White Female) = E(Earnings/Black Male) – E(Earnings/White Male) = β B .

12. For the earnings equation given in (4.28), how would you set up the F -test and what are the restricted and unrestricted regressions for testing the following hypotheses: (a) The equality of slopes and intercepts for Males and Females. (b) The equality of intercepts given the same slopes for Males and Females. Show that the resulting F -statistic is the square of a t-statistic from the unrestricted regression. (c) The equality of intercepts allowing for different slopes for Males and Females. Show that the resulting F -statistic is the square of a t-statistic from the unrestricted regression. (d) Apply your results in parts (a), (b) and (c) to the earnings data underlying Table 4.1. 13. For the earnings data regression underlying Table 4.1. (a) Replicate the regression results given in Table 4.1. (b) Verify that the joint significance of all slope coefficients can be obtained from (4.20). (c) How would you test the joint restriction that expected earnings are the same for Males and Females whether Black or Non-Black holding everything else constant? (d) How would you test the joint restriction that expected earnings are the same whether the individual is married or not and whether this individual belongs to a Union or not? (e) From Table 4.1 what is your estimate of the % change in earnings due to Union membership? If the disturbances are assumed to be log-normal, what would be the estimate suggested by Kennedy (1981) for this % change in earnings? (f) What is your estimate of the % change in earnings due to the individual being married? 14. Crude Quality. Using the data set of U.S. oil field postings on crude prices ($/barrel), gravity (degree API) and sulphur (% sulphur) given in the CRUDES.ASC file on the Springer web site. (a) Estimate the following multiple regression model: POIL = β 1 +β 2 GRAVITY + β 3 SULPHUR + ǫ. (b) Regress GRAVITY = α0 + α1 SULPHUR + ν t then compute the residuals ( ν t ). Now perform the regression P OIL = γ 1 + γ 2  νt + ǫ

 in part (a). What does this tell you? Verify that γ 2 is the same as β 2

(c) Regress POIL = φ1 + φ2 SULPHUR + w. Compute the residuals (w).  Now regress w  on  ν     obtained from part (b), to get w t = δ 1 + δ 2  ν t + residuals. Show that δ 2 = β 2 in part (a). Again, what does this tell you?

(d) To illustrate how additional data affects multicollinearity, show how your regression in part (a) changes when the sample is restricted to the first 25 crudes. (e) Delete all crudes with sulphur content outside the range of 1 to 2 percent and run the multiple regression in part (a). Discuss and interpret these results. 15. Consider the U.S. gasoline data from 1950-1987 given in Table 4.2, and obtained from the file USGAS.ASC on the Springer web site.


Problems Table 4.2 U.S. Gasoline Data: 1950–1987

Year

CAR

QMG (1,000 Gallons)

PMG ($ )

POP (1,000)

RGNP (Billion)

PGNP

1950 1951 1952 1953 1954 1955 1956 1957 1958 1959 1960 1961 1962 1963 1964 1965 1966 1967 1968 1969 1970 1971 1972 1973 1974 1975 1976 1977 1978 1979 1980 1981 1982 1983 1984 1985 1986 1987

49195212 51948796 53301329 56313281 58622547 62688792 65153810 67124904 68296594 71354420 73868682 75958215 79173329 82713717 86301207 90360721 93962030 96930949 101039113 103562018 106807629 111297459 117051638 123811741 127951254 130918918 136333934 141523197 146484336 149422205 153357876 155907473 156993694 161017926 163432944 168743817 173255850 177922000

40617285 43896887 46428148 49374047 51107135 54333255 56022406 57415622 59154330 61596548 62811854 63978489 62531373 64779104 67663848 70337126 73638812 76139326 80772657 85416084 88684050 92194620 95348904 99804600 100212210 102327750 106972740 110023410 113625960 107831220 100856070 100994040 100242870 101515260 102603690 104719230 107831220 110467980

0.272 0.276 0.287 0.290 0.291 0.299 0.310 0.304 0.305 0.311 0.308 0.306 0.304 0.304 0.312 0.321 0.332 0.337 0.348 0.357 0.364 0.361 0.388 0.524 0.572 0.595 0.631 0.657 0.678 0.857 1.191 1.311 1.222 1.157 1.129 1.115 0.857 0.897

152271 154878 157553 160184 163026 165931 168903 171984 174882 177830 180671 183691 186538 189242 191889 194303 196560 198712 200706 202677 205052 207661 209896 211909 213854 215973 218035 220239 222585 225055 227757 230138 232520 234799 237001 239279 241613 243915

1090.4 1179.2 1226.1 1282.1 1252.1 1356.7 1383.5 1410.2 1384.7 1481.0 1517.2 1547.9 1647.9 1711.6 1806.9 1918.5 2048.9 2100.3 2195.4 2260.7 2250.7 2332.0 2465.5 2602.8 2564.2 2530.9 2680.5 2822.4 3115.2 3192.4 3187.1 3248.8 3166.0 3279.1 3489.9 3585.2 3676.5 3847.0

26.1 27.9 28.3 28.5 29.0 29.3 30.3 31.4 32.1 32.6 33.2 33.6 34.0 34.5 35.0 35.7 36.6 37.8 39.4 41.2 43.4 45.6 47.5 50.2 55.1 60.4 63.5 67.3 72.2 78.6 85.7 94.0 100.0 103.9 107.9 111.5 114.5 117.7

CAR: RMG: PMG:

Stock of Cars Motor Gasoline Consumption Retail Price of Motor Gasoline

POP: Population RGNP: Real GNP in 1982 dollars PGNP: GNP Deflator (1982=100)

89


90

Chapter 4: Multiple Regression Analysis (a) For the period 1950-1972 estimate models (1) and (2): logQM G =

β 1 + β 2 logCAR + β 3 logP OP + β 4 logRGN P

(1)

+β 5 logP GN P + β 6 logP M G + u log

RGN P CAR PMG QM G = γ 1 + γ 2 log + γ 3 log + γ 4 log +ν CAR P OP P OP P GN P

(2)

(b) What restrictions should the β’s satisfy in model (1) in order to yield the γ’s in model (2)? (c) Compare the estimates and the corresponding standard errors from models (1) and (2). (d) Compute the simple correlations among the X’s in model (1). What do you observe? (e) Use the Chow-F test to test the parametric restrictions obtained in part (b). (f) Estimate equations (1) and (2) now using the full data set 1950-1987. Discuss briefly the effects on individual parameter estimates and their standard errors of the larger data set. (g) Using a dummy variable, test the hypothesis that gasoline demand per CAR permanently shifted downward for model (2) following the Arab Oil Embargo in 1973? (h) Construct a dummy variable regression that will test whether the price elasticity has changed after 1973. 16. Consider the following model for the demand for natural gas by residential sector, call it model (1): logConsit = β 0 + β 1 logP git + β 2 logP oit + β 3 logP eit + β 4 logHDDit + β 5 logP Iit + uit where i = 1, 2, . . . , 6 states and t = 1, 2, . . . , 23 years. Cons is the consumption of natural gas by residential sector, P g, P o and P e are the prices of natural gas, distillate fuel oil, and electricity of the residential sector. HDD is heating degree days and P I is real per capita personal income. The data covers 6 states: NY, FL, MI, TX, UT and CA over the period 1967-1989. It is given in the NATURAL.ASC file on the Springer web site. (a) Estimate the above model by OLS. Call this model (1). What do the parameter estimates imply about the relationship between the fuels? (b) Plot actual consumption versus the predicted values. What do you observe? (c) Add a dummy variable for each state except California and run OLS. Call this model (2). Compute the parameter estimates and standard errors and compare to model (1). Do any of the interpretations of the price coefficients change? What is the interpretation of the New York dummy variable? What is the predicted consumption of natural gas for New York in 1989? (d) Test the hypothesis that the intercepts of New York and California are the same. (e) Test the hypothesis that all the states have the same intercept. (f) Add a dummy variable for each state and run OLS without an intercept. Call this model (3). Compare the parameter estimates and standard errors to the first two models. What is the interpretation of the coefficient of the New York dummy variable? What is the predicted consumption of natural gas for New York in 1989? (g) Using the regression in part (f), test the hypothesis that the intercepts of New York and California are the same.


References

91

References This chapter draws upon the material in Kelejian and Oates (1989) and Wallace and Silver (1988). Several econometrics books have an excellent discussion on dummy variables, see Gujarati (1978), Judge et al. (1985), Kennedy (1992), Johnston (1984) and Maddala (2001), to mention a few. Other readings referenced in this chapter include: Baltagi, B.H. (1987a), “To Pool or Not to Pool: The Quality Bank Case,” The American Statistician, 41: 150-152. Baltagi, B.H. (1987b), “Simple versus Multiple Regression Coefficients,” Econometric Theory, Problem 87.1.1, 3: 159. Chow, G.C. (1960), “Tests of Equality Between Sets of Coefficients in Two Linear Regressions,” Econometrica, 28: 591-605. Cornwell, C. and P. Rupert (1988), “Efficient Estimation with Panel Data: An Empirical Comparison of Instrumental Variables Estimators,” Journal of Applied Econometrics, 3: 149-155. Dufour, J.M. (1980), “Dummy Variables and Predictive Tests for Structural Change,” Economics Letters, 6: 241-247. Dufour, J.M. (1982), “Recursive Stability of Linear Regression Relationships,” Journal of Econometrics, 19: 31-76. Gujarati, D. (1970), “Use of Dummy Variables in Testing for Equality Between Sets of Coefficients in Two Linear Regressions: A Note,” The American Statistician, 24: 18-21. Gujarati, D. (1970), “Use of Dummy Variables in Testing for Equality Between Sets of Coefficients in Two Linear Regressions: A Generalization,” The American Statistician, 24: 50-52. Halvorsen, R. and R. Palmquist (1980), “The Interpretation of Dummy Variables in Semilogarithmic Equations,” American Economic Review, 70: 474-475. Hanushek, E.A. and J.E. Jackson (1977), Statistical Methods for Social Scientists (Academic Press: New York). Hill, R. Carter and L.C. Adkins (2001), “Collinearity,” Chapter 12 in B.H. Baltagi (ed.) A Companion to Theoretical Econometrics (Blackwell: Massachusetts). Kennedy, P.E. (1981), “Estimation with Correctly Interpreted Dummy Variables in Semilogarithmic Equations,” American Economic Review, 71: 802. Kennedy, P.E. (1981), “The Balentine: A Graphical Aid for Econometrics,” Australian Economic Papers, 20: 414-416. Kennedy, P.E. (1986), “Interpreting Dummy Variables,” Review of Economics and Statistics, 68: 174-175. Nieswiadomy, M. (1986), “Effect of an Additional Regressor on R2 ,” Econometric Theory, Problem 86.3.1, 2:442. Poirier, D. (1976), The Econometrics of Structural Change(North Holland: Amsterdam). Salkever, D. (1976), “The Use of Dummy Variables to Compute Predictions, Prediction Errors, and Confidence Intervals,” Journal of Econometrics, 4: 393-397. Suits, D. (1984), “Dummy Variables: Mechanics vs Interpretation,” Review of Economics and Statistics, 66: 132-139.


92

Chapter 4: Multiple Regression Analysis

Suits, D.B., A. Mason and L. Chan (1978), “Spline Functions Fitted by Standard Regression Methods,” Review of Economics and Statistics, 60: 132-139. Utts, J. (1982), “The Rainbow Test for Lack of Fit in Regression,”Communications in Statistics-Theory and Methods, 11: 1801-1815.

Appendix Residual Interpretation of Multiple Regression Estimates Proof of Claim 1: Regressing X2 on all the other X’s yields residuals  ν 2 that satisfy the usual properties of OLS residuals similar to those in (4.2), i.e., n n n n ν 2i XKi = 0 (A.1) ν 2i X4i = .. = i=1  ν 2i X3i = i=1  2i = 0, i=1  i=1 ν  Note that X2 is the dependent nvariable of this regression, and X2 is the predicted value from this 2i = 0. This holds because X 2 is a linear combination of the regression. The latter satisfies i=1 ν2i X other X’s, all of which satisfy (A.1). Turn now to the estimated regression equation:  X2i + .. + β  XKi + ei Yi = α +β 2 K

(A.2)

Multiply (A.2) by X2i and sum n

i=1

X2i Yi = α 

n

i=1

n

 n X2i XKi  n X 2 + .. + β X2i + β K 2 2i i=1 i=1

(A.3)

This uses the fact that i=1 X2i ei = 0. Alternatively, (A.3) is just the second equation from (4.3). 2i +  Substituting X2i = X ν 2i , in (A.3) one gets n

i=1

 2i Yi + n  2i + β  2 + ..  n X X ν 2i Yi = α  ni=1 X 2 2i i=1 i=1  n n 2    + β K i=1 X2i XKi + β 2 i=1  ν 2i

2i  2i and sum, we get using (A.1) and the fact that Σni=1 X ν 2i = 0. Multiply (A.2) by X n

i=1

n

2i Yi = α  X

n

  n X  n X  2i XKi + n X 2i X2i + .. + β 2i + β X K 2 i=1 2i ei i=1 i=1

n

2  n X  n X  2i 2i + β + .. + β X K 2 i=1 2i XKi i=1

i=1

(A.4)

(A.5)

  linear combination of all the other X’s, all of which satisfy (4.2). Also, i=1 X2 ei = 0 since X2 is a 2i   2 since n X 2i X2i = n X X ν 2i = 0. Hence (A.5) reduces to i=1 i=1 2i i=1

But n

n

i=1

2i Yi = α X 

i=1

(A.6)

Subtracting (A.6) from (A.4), we get n

i=1

 ν2i Yi = β 2

n

i=1

ν22i

 is the slope estimate of the simple regression of Y on  and β ν 2 as given in (4.5). 2 By substituting for Yi its expression from equation (4.1) in (4.5) we get n n n  = β n X2i  β ν 22i ν 2i ui / i=1  ν 22i + i=1  ν 2i / i=1  2 2 i=1

(A.7)

(A.8)

n  n n 2i +  ν 2i = 0, which implies ν 2i = 0. But, X2i = X ν 2i and i=1 X ν 2i = 0 and i=1  where i=1 X1i  2i    n n n  =β + n   is unbiased with ν 22i and β ν 22i . This means that β ν 2i = i=1  that i=1 X2i  2 2 2 i=1 ν 2i ui / i=1 


Appendix

93

 ) = β since  E(β ν 2 is a linear combination of the X’s and these in turn are not correlated with the u’s. 2 2 Also, n n  − β )2 = E(n   ) = E(β ν 22i )2 = σ 2 / i=1  ν 22i var(β 2 2 2 i=1 ν 2i ui / i=1 

 for k = 2, . . . , K, i.e., The same results apply for any β k   n n 2  ν ki ν ki Yi / i=1  β k = i=1 

(A.9)

where  ν k is the OLS residual of Xk on all the other X’s in the regression. Similarly,    β k = β k + ni=1  ν ki ui / ni=1  ν 2ki (A.10)  n 2  ) = σ2 /  ) = β with var(β and E(β ν ki for k = 2, . . . , K. Note also that k k k i=1  n n n  ,β  ) = E(β  − β )(β  − β ) = E(n  cov(β ν 2ki ) ν 22i )( i=1 νki ui / i=1  2 k 2 2 k k i=1  i=1 ν 2i ui /    n n n 2 2 ν ki ν 2i i=1  ν 2i  ν ki / i=1  = σ 2 i=1 

Proof of Claim 2: Regressing Y on all the other X’s yields, Yi = Yi + ω  i . Substituting this expression for Yi in (4.5) one gets n n n  = (n   n  ν 22i (A.11) ν 2i ω  i / i=1  ν 22i = i=1   i )/ i=1  β 2 i=1 ν 2i ω i=1 ν 2i Yi + where the last equality follows from the fact that Y is a linear combination of all X’s excluding X2 , all  is the estimate of the slope coefficient in the linear regression of ω  on of which satisfy (A.1). Hence β 2 ν2. 

Simple, Partial and Multiple Correlation Coefficients

2 , as the proportion of In Chapter 3, we interpreted the square of the simple correlation coefficient, rY,X 2 2 the variation in Y that is explained by X2 . Similarly, rY,Xk is the R-squared of the simple regression of Y on Xk for k = 2, . . . , K. In fact, one can compute these simple correlation coefficients and find out which Xk is most correlated with Y , say it is X2 . If one is selecting regressors to include in the regression equation, X2 would be the best one variable candidate. In order to determine what variable to include next, we look at partial correlation coefficients of the form rY,Xk .X2 for k = 2. The square of this first-order partial gives the proportion of the residual variation in Y , not explained by X2 , that is explained by the addition of Xk . The maximum first-order partial (‘first’ because it has only one variable after the dot) determines the best candidate to follow X2 . Let us assume it is X3 . The first-order partial correlation coefficients can be computed from simple correlation coefficients as follows: rY,X3 − rY,X2 rX2 ,X3  rY,X3 .X2 =  2 2 1 − rY,X 1 − rX 2 2 ,X3

see Johnston (1984). Next we look at second-order partials of the form rY,Xk .X2 ,X3 for k = 2, 3, and so on. This method of selecting regressors is called forward selection. Suppose there is only X2 , X3and X4 2 ) is the proportion of the variation in Y , i.e., ni=1 yi2 , in the regression equation. In this case (1 − rY,X 2 2 2 that is not explained by X2 . Also (1 − rY,X3 .X2 )(1 − rY,X ) denotes the proportion of the variation in Y 2 2 2 2 ) )(1 − rY,X )(1 − rY,X not explained after the inclusion of both X2 and X3 . Similarly (1 − rY,X 2 3 .X2 4 .X2 ,X3 is the proportion of the variation in Y unexplained after the inclusion of X2 , X3 and X4 . But this is exactly (1 − R2 ), where R2 denotes the R-squared of the multiple regression of Y on a constant, X2 , X3 2 . Hence and X4 . This R2 is called the multiple correlation coefficient, and is also written as RY.X 2 ,X3 ,X4 2 2 2 2 ) = (1 − rY,X )(1 − rY,X )(1 − rY,X ) (1 − RY.X 2 ,X3 ,X4 2 3 .X2 4 .X2 ,X3

and similar expressions relating the multiple correlation coefficient to simple and partial correlation coefficients can be written by including say X3 first then X4 and X2 in that order.


CHAPTER 5

Violations of the Classical Assumptions 5.1

Introduction

In this chapter, we relax the assumptions made in Chapter 3 one by one and study the effect of that on the OLS estimator. In case the OLS estimator is no longer a viable estimator, we derive an alternative estimator and propose some tests that will allow us to check whether this assumption is violated.

5.2

The Zero Mean Assumption

Violation of assumption 1 implies that the mean of the disturbances is no longer zero. Two cases are considered:

Case 1: E(ui ) = μ = 0 The disturbances have a common mean which is not zero. In this case, one can subtract μ from the ui ’s and get new disturbances u∗i = ui − μ which have zero mean and satisfy all the other assumptions imposed on the ui ’s. Having subtracted μ from ui we add it to the constant α leaving the regression equation intact: Yi = α∗ + βXi + u∗i

i = 1, 2, . . . , n

(5.1)

where α∗ = α + μ. It is clear that only α∗ and β can be estimated, and not α nor μ. In other words, one cannot retrieve α and μ from an estimate of α∗ without additional assumptions or further information, see problem 10. With this reparameterization, equation (5.1) satisfies the four classical assumptions, and therefore OLS gives the BLUE estimators of α∗ and β. Hence, a constant non-zero mean for the disturbances affects only the intercept estimate but not the slope. Fortunately, in most economic applications, it is the slope coefficients that are of interest and not the intercept.

Case 2: E(ui ) = μi The disturbances have a mean which varies with every observation. In this case, one can transform the regression equation as in (5.1) by adding and subtracting μi . The problem, however, is that α∗i = α + μi now varies with each observation, and hence we have more parameters than observations. In fact, there are n intercepts and one slope to be estimated with n observations. Unless we have repeated observations like in panel data, see Chapter 12 or we have some prior information on these α∗i , we cannot estimate this model.


96

Chapter 5: Violations of the Classical Assumptions

5.3

Stochastic Explanatory Variables

Sections 5.5 and 5.6 will study violations of assumptions 2 and 3 in detail. This section deals with violations of assumption 4 and its effect on the properties of the OLS estimators. In this case, X is a random variable which may be (i) independent; (ii) contemporaneously uncorrelated; or (iii) simply correlated with the disturbances. Case 1: If X is independent of u, then all the results of Chapter 3 still hold, but now they are conditional on the particular set of X’s drawn in the sample. To illustrate this result, recall that for the simple linear regression: OLS = β + n wi ui where wi = xi / n x2 (5.2) β i=1 i=1 i n n Hence, when we take expectations E( i=1 wi ui ) = i=1 E(wi )E(ui ) = 0. The first equality holds because X and u are independent and the second equality holds because the u’s have zero mean. In other words the unbiasedness property of the OLS estimator still holds. However, the n n n 2 2 n 2  var(β OLS ) = E( i=1 wi ui ) = i=1 E(wi ) j=1 E(wi wj )E(ui uj ) = σ i=1 where the last equality follows from assumptions 2 and 3, homoskedasticity and no serial correlation. The only difference between this result and that of Chapter 3 is that we have expectations on the X’s rather than the X’s themselves. Hence, by conditioning on the particular set of X’s that are observed, we can use all the results of Chapter 3. Also, maximizing the likelihood involves both the X’s and the u’s. But, as long as the distribution of the X’s does not involve the parameters we are estimating, i.e., α, β and σ 2 , the same maximum likelihood estimators are obtained. Why? Because f (x1 , x2 , . . . , xn , u1 , u2 , . . . , un ) = f1 (x1 , x2 , . . . , xn )f2 (u1 , u2 , . . . , un ) since the X’s and the u’s are independent. Maximizing f with respect to (α, β, σ 2 ) is the same as maximizing f2 with respect to (α, β, σ 2 ) as long as f1 is not a function of these parameters. Case 2: Consider a simple model of consumption, where Yt , current consumption, is a function of Yt−1 , consumption in the previous period. This is the case for a habit forming consumption good like cigarette smoking. In this case our regression equation becomes Yt = α + βYt−1 + ut

t = 2, . . . , T

(5.3)

where we lost one observation due to lagging. It is obvious that Yt is correlated to ut , but the question here is whether Yt−1 is correlated to ut . After all, Yt−1 is our explanatory variable Xt . As long as assumption 3 is not violated, i.e., the u’s are not correlated across periods, ut represents a freshly drawn disturbance independent of previous disturbances and hence is not correlated with the already predetermined Yt−1 . This is what we mean by contemporaneously uncorrelated, i.e., ut is correlated with Yt , but it is not correlated with Yt−1 . The OLS estimator of β is T T T T 2 2  β (5.4) OLS = t=2 yt yt−1 / t=2 yt−1 = β + t=2 yt−1 ut / t=2 yt−1

and the expected value of (5.4) is not β because in general,     2 2 ) = E( Tt=2 yt−1 ut )/E( Tt=2 yt−1 ). E( Tt=2 yt−1 ut / Tt=2 yt−1

The expected value of a ratio is not the ratio of expected values. Also, even if E(Yt−1 ut ) = 0, one can easily show that E(yt−1 ut ) = 0. In fact, yt−1 = Yt−1 − Y¯ , and Y¯ contains Yt in it, and


5.3

Stochastic Explanatory Variables

97

we know that E(Yt ut ) = 0. Hence, we lost the unbiasedness property of OLS. However, all the  asymptotic properties still hold. In fact, β OLS is consistent because  plim β OLS = β + cov(Yt−1 , ut )/var(Yt−1 ) = β

(5.5)

 where the second equality follows from (5.4) and the fact that plim( Tt=2 yt−1 ut /T ) is T 2 /T ) = var(Y cov(Yt−1 , ut ) which is zero, and plim( t=2 yt−1 t−1 ) which is positive and finite.

Case 3: X and u are correlated, in  this case OLS is biased and inconsistent. This  can be easily deduced from (5.2) since plim( ni=1 xi ui /n) is the cov(X, u) = 0, and plim( ni=1 x2i /n) is positive and finite. This means that OLS is no longer a viable estimator, and an alternative estimator that corrects for this bias has to be derived. In fact we will study three specific cases where this assumption is violated. These are: (i) the errors in measurement case; (ii) the case of a lagged dependent variable with correlated errors; and (iii) simultaneous equations. Briefly, the errors in measurement case involves a situation where the true regression model is in terms of X ∗ , but X ∗ is measured with error, i.e., Xi = Xi∗ + ν i , so we observe Xi but not Xi∗ . Hence, when we substitute this Xi for Xi∗ in the regression equation, we get Yi = α + βXi∗ + ui = α + βXi + (ui − βν i )

(5.6)

where the composite error term is now correlated with Xi because Xi is correlated with ν i . After all, Xi = Xi∗ + ν i and E(Xi ν i ) = E(ν 2i ) if Xi∗ and ν i are uncorrelated. Similarly, in case (ii) above, if the u’s were correlated across time, i.e., ut−1 is correlated with ut , then Yt−1 , which is a function of ut−1 , will also be correlated with ut , and E(Yt−1 ut ) = 0. More on this and how to test for serial correlation in the presence of a lagged dependent variable in Chapter 6. Finally, if one considers a demand and supply equations where quantity Qt is a function of price Pt in both equations Qt = α + βPt + ut

(demand)

(5.7)

Qt = δ + γPt + ν t

(supply)

(5.8)

The question here is whether Pt is correlated with the disturbances ut and ν t in both equations. The answer is yes, because (5.7) and (5.8) are two equations in two unknowns Pt and Qt . Solving for these variables, one gets Pt as well as Qt as a function of a constant and both ut and ν t . This means that E(Pt ut ) = 0 and E(Pt ν t ) = 0 and OLS performed on either (5.7) or (5.8) is biased and inconsistent. We will study this simultaneous bias problem more rigorously in Chapter 11. For all situations where X and u are correlated, it would be illuminating to show graphically why OLS is no longer a consistent estimator. Let us consider the case where the disturbances are, say, positively correlated with the explanatory variable. Figure 3.3 of Chapter 3 shows the true regression line α + βXi . It also shows that when Xi and ui are positively correlated then an Xi higher than its mean will be associated with a disturbance ui above its mean, i.e., a positive disturbance. Hence, Yi = α + βXi + ui will always be above the true regression line whenever Xi is above its mean. Similarly Yi would be below the true regression line for every Xi below its mean. This means that not knowing the true regression line, a researcher fitting OLS on this data will have a biased intercept and slope. In fact, the intercept will be understated and the slope will be overstated. Furthermore, this bias does not disappear with more data, since


98

Chapter 5: Violations of the Classical Assumptions

this new data will be generated by the same mechanism described above. Hence these OLS estimates are inconsistent. Similarly, if Xi and ui are negatively correlated, the intercept will be overstated and the slope will be understated. This story applies to any equation with at least one of its right hand side variables correlated with the disturbance term. Correlation due to the lagged dependent variable with autocorrelated errors, is studied in Chapter 6, whereas the correlation due to the simultaneous equations problem is studied in Chapter 11.

5.4

Normality of the Disturbances

If the disturbance are not normal, OLS is still BLUE provided assumptions 1-4 still hold. Normality made the OLS estimators minimum variance unbiased MVU and these OLS estimators turn out to be identical to the MLE. Normality allowed the derivation of the distribution of these estimators and this in turn allowed testing of hypotheses using the t and F -tests considered in the previous chapter. If the disturbances are not normal, yet the sample size is large, one can still use the normal distribution for the OLS estimates asymptotically by relying on the Central Limit Theorem, see Theil (1978). Theil’s proof is for the case of fixed X’s in repeated samples, zero mean and constant variance on the disturbances. A simple asymptotic test for the normality assumption is given by Jarque and Bera (1987). This is based on the fact that the normal distribution has a skewness measure of zero and a kurtosis of 3. Skewness (or lack of symmetry) is measured by S=

Square of the 3rd moment about the mean [E(X − μ)3 ]2 = [E(X − μ)2 ]3 Cube of the variance

Kurtosis (a measure of flatness) is measured by κ=

E(X − μ)4 4th moment about the mean = [E(X − μ)2 ]2 Square of the variance

For the normal distribution S = 0 and κ = 3. Hence, the Jarque-Bera (JB) statistic is given by

2 (κ − 3)2 S JB = n + 6 24 where S represents skewness and κ represents kurtosis of the OLS residuals. This statistic is asymptotically distributed as χ2 with two degrees of freedom under H0 . Rejecting H0 , rejects normality of the disturbances but does not offer an alternative distribution. In this sense, the test is non-constructive. In addition, not rejecting H0 does not necessarily mean that the distribution of the disturbances is normal, it only means we do not reject that the distribution of the disturbances is symmetric and has a kurtosis of 3. See the empirical example in section 5.5 for an illustration. The Jarque-Bera test is part of the standard output using EViews.

5.5

Heteroskedasticity

Violation of assumption 2, means that the disturbances have a varying variance, i.e., E(u2i ) = σ 2i , i = 1, 2, . . . , n. First, we study the effect of this violation on the OLS estimators. For the simple


5.5

Heteroskedasticity

99

 OLS given in equation (5.2) is still unbiased and consistent linear regression it is obvious that β because these properties depend upon assumptions 1 and 4, and not assumption 2. However,  the variance of β OLS is now different n n n 2 2 2 2 n 2 2  var(β (5.9) OLS ) = var( i=1 wi ui ) = i=1 wi σ i = i=1 xi σ i /( i=1 xi )

where the second equality follows from assumption 3 and the fact that var(ui ) is now σ 2i .   Note that if σ 2i = σ 2 , this reverts back to σ 2 / ni=1 x2i , the usual formula for var(β OLS ) under 2 2 homoskedasticity. Furthermore, one can show that E(s ) will involve all of the σ i ’s and not one common σ 2 , see problem 1. This means that the regression package reporting s2 / ni=1 x2i as the  estimate of the variance of β OLS is committing two errors. One, it is not using the right formula for the variance, i.e., equation (5.9). Second, it is using s2 to estimate a common σ 2 when in  n  fact the σ 2i ’s are different. The bias from using s2 / i=1 x2i as an estimate of var(β OLS ) will 2 depend upon the nature of the heteroskedasticity and the regressor. In fact, if σ i is positively  related to x2i , one can show that s2 / ni=1 x2i understates the true variance and hence the tstatistic reported for β = 0 is overblown, and the confidence interval for β is tighter than it is supposed to be, see problem 2. This means that the t-statistic in this case is biased towards rejecting H0 ; β = 0, i.e., showing significance of the regression slope coefficient, when it may not be significant. The OLS estimator of β is linear unbiased and consistent, but is it still BLUE? In order to answer this question, we note that the only violation we have is that the var(ui ) = σ 2i . Hence, if we divided ui by σ i /σ, the resulting u∗i = σui /σ i will have a constant variance σ 2 . It is easy to show that u∗ satisfies all the classical assumptions including homoskedasticity. The regression model becomes σYi /σ i = ασ/σ i + βσXi /σ i + u∗i

(5.10)

and OLS on this model (5.10) is BLUE. The OLS normal equations on (5.10) are n n n 2 2 2 i=1 (Yi /σ i ) = α i=1 (1/σ i ) + β i=1 (Xi /σ i ) n

2 i=1 (Yi Xi /σ i ) = α

n

2 i=1 (Xi /σ i ) + β

n

(5.11)

2 2 i=1 (Xi /σ i )

Note that σ 2 drops out of these equations. Solving (5.11), see problem 3, one gets    n (Xi /σ 2 )/ n (1/σ 2 )] = Y¯ ∗ − β X ¯∗ α  = [ ni=1 (Y1i /σ 2i )/ ni=1 (1/σ 2i )] − β[ i i i=1 i=1     with Y¯ ∗ = [ ni=1 (Yi /σ 2i )/ ni=1 (1/σ 2i )] = ni=1 wi∗ Yi / ni=1 wi∗ and ¯ ∗ = [n (Xi /σ 2 )/ n (1/σ 2 )] = n w∗ Xi / n w∗ X i i i=1 i=1 i=1 i i=1 i where wi∗ = (1/σ 2i ). Similarly,     [ ni=1 (1/σ 2i )][ ni=1 (Yi Xi /σ 2i )] − [ ni=1 (Xi /σ 2i )][ ni=1 (Yi /σ 2i )]     β = [ ni=1 Xi2 /σ 2i ][ ni=1 (1/σ 2i )] − [ ni=1 (Xi /σ 2i )]2     ( ni=1 wi∗ )( ni=1 wi∗ Xi Yi ) − ( ni=1 wi∗ Xi )( ni=1 wi∗ Yi )    = ( ni=1 wi∗ )( ni=1 wi∗ Xi2 ) − ( ni=1 wi∗ Xi )2 n ¯ ∗ )(Yi − Y¯ ∗ ) w∗ (X − X i=1 ni i∗ = ¯∗ 2 i=1 wi (Xi − X )

(5.12a)

(5.12b)


100

Chapter 5: Violations of the Classical Assumptions

 obtained from the regression in (5.10), are different It is clear that the BLU estimators α  and β, 2  from the usual OLS estimators α  OLS and β OLS since they depend upon the σ i ’s. It is also true 2 2 that when σ i = σ for all i = 1, 2, . . . , n, i.e., under homoskedasticity, (5.12) reduces to the usual OLS estimators given by equation (3.4) of Chapter 3. The BLU estimators weight the i-th observation by (1/σ i ) which is a measure of precision of that observation. The more precise the observation, i.e., the smaller σ i , the larger is the weight attached to that observation. α   are also known as Weighted Least Squares (WLS) estimators which are a specific form and β of Generalized Least Squares (GLS). We will study GLS in details in Chapter 9, using matrix notation. Under heteroskedasticity, OLS looses efficiency in that it is no longer BLUE. However, because it is still unbiased and consistent and because the true σ 2i ’s are never known some researchers compute OLS as an initial consistent estimator of the regression coefficients. It is important to emphasize however, that the standard errors of these estimates as reported by the regression package are biased and any inference based on these estimated variances including the reported t-statistics are misleading. White (1980) proposed a simple procedure that would yield heteroskedasticity consistent standard errors of the OLS estimators. In equation (5.9), this amounts to replacing σ 2i by e2i , the square of the i-th OLS residual, i.e.,  W hite’s var(β OLS ) =

n

2 2 n 2 2 i=1 xi ei /( i=1 xi )

(5.13)

Note that we can not consistently estimate σ 2i by e2i , since there is one observation per parameter estimated. As the sample size increases, so does the number of unknown σ 2i ’s. What White (1980) 2  consistently estimates is the var(β OLS ) which is a weighted average of the ei . The same analysis applies to the multiple regression OLS estimates. In this case, White’s (1980) heteroskedasticity consistent estimate of the variance of the k-th OLS regression coefficient β k , is given by 2 2 n  ) = n  2ki )2 W hite’s var(β k i=1 ν ki ei /( i=1 ν

where  ν 2k is the squared OLS residual obtained from regressing Xk on the remaining regressors in the equation being estimated. ei is the i-th OLS residual from this multiple regression equation. Many regression packages provide White’s heteroskedasticity-consistent estimates of the variances and their corresponding robust t-statistics. For example, using EViews, one clicks on Quick, choose Estimate Equation. Now click on Options, a menu appears where one selects White to obtain the heteroskedasticity-consistent estimates of the variances. While the regression packages correct for heteroskedasticity in the t-statistics they do not usually do that for the F -statistics studied, say in Example 2 in Chapter 4. Wooldridge (1991) suggests a simple way of obtaining a robust LM statistic for H0 ; β 2 = β 3 = 0 in the multiple regression (4.1). This involves the following steps: (1) Run OLS on the restricted model without X2 and X3 and obtain the restricted least squares residuals u . (2) Regress each of the independent variables excluded under the null (i.e., X2 and X3 ) on all of the other included independent variables (i.e., X4 , X5 , . . . , XK ) including the constant. Get the corresponding residuals v2 and v3 , respectively.

(3) Regress the dependent variable equal to 1 for all observations on v2 u , v3 u  without a constant and obtain the robust LM statistic equal to the n- the sum of squared residuals of


5.5

Heteroskedasticity

101

this regression. This is exactly nRu2 of this last regression. Under H0 this LM statistic is distributed as χ22 .  The only problem is that the σ i ’s Since OLS is no longer BLUE, one should compute α  and β. are rarely known. One example where the σ i ’s are known up to a scalar constant is the following simple example of aggregation. Example 5.1: Aggregation and Heteroskedasticity. Let Yij be the observation on the j-th firm in the i-th industry, and consider the following regression: Yij = α + βXij + uij

j = 1, 2, . . . , ni ; i = 1, 2, . . . , m

(5.14)

If only aggregate observations on each industry are available, then (5.14) is summed over firms, i.e., (5.15) Yi = αni + βXi + ui i = 1, 2, . . . , m ni ni ni where Yi = j=1 Yij , Xi = j=1 Xij , ui = j=1 uij for i = 1, 2, . . . , m. Note that although the uij ’s are IID(0, σ 2 ), by aggregating, we get ui ∼ (0, ni σ 2 ). This means that the disturbances in (5.15) are heteroskedastic. However, σ 2i = ni σ 2 and is known up to a scalar constant. In fact, σ/σ i is 1/(ni )1/2 . Therefore, premultiplying (5.15) by 1/(ni )1/2 and performing OLS on the transformed equation results in BLU estimators of α and β. In other words, BLU estimation reduces to performing OLS of Yi /(ni )1/2 on (ni )1/2 and Xi /(ni )1/2 , without an intercept. There may be other special cases in practice where σ i is known up to a scalar, but in general, σ i is usually unknown and will have to be estimated. This is hopeless with only n observations, since there are n σ i ’s, so we either have to have repeated observations, or know more about the σ i ’s. Let us discuss these two cases. Case 1: Repeated Observations Suppose that ni households are selected randomly with income Xi for i = 1, 2, . . . , m. For each household j = 1, 2, . . . , ni , we observe its consumption expenditures on food, say Yij . The regression equation is Yij = α + βXi + uij

i = 1, 2, . . . , m ; j = 1, 2, . . . , ni

(5.16)

where m is the number of income groups selected. Note that Xi has only one subscript, whereas Yij has two subscripts denoting the repeated observations on households with the same income Xi . The uij ’s are independently distributed (0, σ 2i ) reflecting the heteroskedasticity in consumption expenditures among the different income groups. In this case, there are n = m i=1 ni observations and m σ 2i ’s to be estimated. This is feasible, and there are two methods for estimating these σ 2i ’s. The first is to compute  i (Yij − Y¯i )2 /(ni − 1) s2i = nj=1

ni ni 2 2i = where Y¯i = j=1 Yij /ni . The second is to compute s j=1 eij /ni where eij is the OLS residual given by  eij = Yij − α  OLS − β OLS Xi


102

Chapter 5: Violations of the Classical Assumptions

Both estimators of σ 2i are consistent. Substituting either s2i or s2i for σ 2i in (5.12) will result in feasible estimators of α and β. However, the resulting estimates are no longer BLUE. The substitution of the consistent estimators of σ 2i is justified on the basis that the resulting α and β estimates will be asymptotically efficient, see Chapter 9. Of course, this step could have si on (1/ si ) and (Xi / si ) without a constant, or the similar been replaced by a regression of Yij / regression in terms of si . For this latter estimate, s2i , one can iterate, i.e., obtaining new residuals based on the new regression estimates and therefore new s2i . The process continues until the estimates obtained from the r-th iteration do not differ from those of the (r + 1)th iteration in absolute value by more than a small arbitrary positive number chosen as the convergence criterion. Once the estimates converge, the final round estimators are the maximum likelihood estimators, see Oberhofer and Kmenta (1974). Case 2: Assuming More Information on the Form of Heteroskedasticity If we do not have repeated observations, it is hopeless to try and estimate n variances and α and β with only n observations. More structure on the form of heteroskedasticity is needed to estimate this model, but not necessarily to test it. Heteroskedasticity is more likely to occur with cross-section data where the observations may be on firms with different size. For example, a regression relating profits to sales might have heteroskedasticity, because larger firms have more resources to draw upon, can borrow more, invest more, and loose or gain more than smaller firms. Therefore, we expect the form of heteroskedasticity to be related to the size of the firm, which is reflected in this case by the regressor, sales, or some other variable that measures size, like assets. Hence, for this regression we can write σ 2i = σ 2 Zi2 , where Zi denotes the sales or assets of firm i. Once again the form of heteroskedasticity is known up to a scalar constant and the BLU estimators of α and β can be obtained from (5.12), assuming Zi is known. Alternatively, one can run the regression of Yi /Zi on 1/Zi and Xi /Zi without a constant to get the same result. Special cases of Zi are Xi and E(Yi ). (i) If Zi = Xi the regression becomes that of Yi /Xi on 1/Xi and a constant. Note that the regression coefficient of 1/Xi is the estimate of α, while the constant of the regression is now the estimate of β. But, is it possible to have ui uncorrelated with Xi when we are assuming var(ui ) related to Xi ? The answer is yes, as long as E(ui /Xi ) = 0, i.e., the mean of ui is zero for every value of Xi , see Figure 3.4 of Chapter 3. This, in turn, implies that the overall mean of the ui ’s is zero, i.e., E(ui ) = 0 and that cov(Xi , ui ) = 0. If the latter is not satisfied and say cov(Xi , ui ) is positive, then large values of Xi imply large values of ui . This would mean that for these values of Xi , we have a nonzero mean for the corresponding ui ’s. This contradicts E(ui /Xi ) = 0. Hence, if E(ui /Xi ) = 0, then cov(Xi , ui ) = 0. (ii) If Zi = E(Yi ) = α + βXi , then σ 2i is proportional to the population regression line, which is a linear function of α and β. Since the OLS estimates are consistent     OLS + β one can estimate E(Yi ) by Yi = α OLS Xi use Zi = Yi instead of E(Yi ). In other words,    run the regression of Yi /Yi on 1/Yi and Xi /Yi without a constant. The resulting estimates are asymptotically efficient, see Amemiya (1973). One can generalize σ 2i = σ 2 Zi2 to σ 2i = σ 2 Ziδ where δ is an unknown parameter to be estimated. Hence rather than estimating n σ 2i ’s one has to estimate only σ 2 and δ. Assuming normality one can set up the likelihood function and derive the first-order conditions by differentiating that likelihood with respect to α, β, σ 2 and δ. The resulting equations are highly nonlinear. Alternatively, one can search over possible values for δ = 0, 0.1, 0.2, . . . , 4, and get the δ/2 δ/2 δ/2 corresponding estimates of α, β, and σ 2 from the regression of Yi /Zi on 1/Zi and Xi /Zi


5.5

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103

without a constant. This is done for every δ and the value of the likelihood function is reported. Using this search procedure one can get the maximum value of the likelihood and corresponding to it the MLE of α, β, σ 2 and δ. Note that as δ increases so does the degree of heteroskedasticity. Problem 4 asks the reader to compute the relative efficiency of the OLS estimator with respect to the BLU estimator for Zi = Xi for various values of δ. As expected the relative efficiency of the OLS estimator declines as the degree of heteroskedasticity increases. One can also generalize σ 2i = σ 2 Ziδ to include more Z variables. In fact, a general form of this multiplicative heteroskedasticity is logσ 2i = logσ 2 + δ 1 logZ1i + δ 2 logZ2i + . . . + δr logZri

(5.17)

with r < n, otherwise one cannot estimate with n observations. Z1 , Z2 , . . . , Zr are known variables determining the heteroskedasticity. Note that if δ 2 = δ 3 = . . . = δ r = 0, we revert back to σ 2i = σ 2 Ziδ , where δ = δ1 . For the estimation of this general multiplicative form of heteroskedasticity, see Harvey (1976). Another form for heteroskedasticity, is the additive form σ 2i = a + b1 Z1i + b2 Z2i + . . . + br Zri

(5.18)

where r < n, see Goldfeld and Quandt (1972). Special cases of (5.18) include σ 2i = a + b1 Xi + b2 Xi2

(5.19)

where if a and b1 are zero we have a simple form of multiplicative heteroskedasticity. In order to estimate the regression model with additive heteroskedasticity of the type given in (5.19), one can get the OLS residuals, the ei ’s, and run the following regression e2i = a + b1 Xi + b2 Xi2 + vi e2i

(5.20)

σ 2i .

where vi = − The vi ’s are heteroskedastic, and the OLS estimates of (5.20) yield the following estimates of σ 2i aOLS + b1,OLS Xi + b2,OLS Xi2 σ 2i = 

(5.21)

(e2i / σ i ) = a(1/ σ i ) + b1 (Xi / σ i ) + b2 (Xi2 / σ i ) + wi

(5.22)

One can obtain a better estimate of the σ 2i ’s by computing the following regression which corrects for the heteroskedasticity in the vi ’s The new estimates of σ 2i are a + b1 Xi + b2 Xi2 σ 2i = 

(5.23)

one can run the regression of where  a,  b1 and b2 are the OLS estimates from (5.22). Using the Yi / σ i on (1/ σ i ) and Xi / σ i without a constant to get asymptotically efficient estimates of α and β. These have the same asymptotic properties as the MLE estimators derived in Rutemiller and Bowers (1968), see Amemiya (1977) and Buse (1984). The problem with this iterative procedure is that there is no guarantee that the σ 2i ’s are positive, which means that the square root σ i may not exist. This problem would not occur if σ 2i = (a + b1 Xi + b2 Xi2 )2 because in this case one regresses |ei | on a constant, Xi and Xi2 and the predicted value from this regression would be an estimate of σ i . It would not matter if this predictor is negative, because we do not have to take its square root and because its sign cancels in the OLS normal equations of the final regression of Yi / σ i on (1/ σ i ) and (Xi / σ i ) without a constant. σ 2i ’s


104

Chapter 5: Violations of the Classical Assumptions

Testing for Homoskedasticity In the repeated observation’s case, one can perform Bartlett’s (1937) test. The null hypothesis is H0 ; σ 21 = σ 22 = . . . = σ 2m . Under σ 2 which can be estimated m the2 null there is onevariance m 2 by the pooled variance s = i /ν where ν = i=1 ν i s i=1 ν i , and ν i = ni − 1. Under the alternative hypothesis there are m different variances estimated by s2i for i = 1, 2, . . . , m. The Likelihood Ratio test, which computes the ratio of the likelihoods under the null and alternative hypotheses, reduces to computing B = [νlogs2 −

m

i=1 ν i

logs2i ]/c

(5.24)

 2 where c = 1 + [ m i=1 (1/ν i ) − 1/ν] /3(m − 1). Under H0 , B is distributed χm−1 . Hence, a large p-value for the B-statistic given in (5.24) means that we do not reject homoskedasticity whereas, a small p-value leads to rejection of H0 in favor of heteroskedasticity. In case of no repeated observations, several tests exist in the literature. Among these are the following: (1) Glejser’s (1969) Test: In this case one regresses |ei | on a constant and Ziδ for δ = 1, −1, 0.5 and −0.5. If the coefficient of Ziδ is significantly different from zero, this would lead to a rejection of homoskedasticity. The power of this test depends upon the true form of heteroskedasticity. One important result however, is that this power is not seriously impaired if the wrong value of δ is chosen, see Ali and Giaccotto (1984) who confirmed this result using extensive Monte Carlo experiments. (2) The Goldfeld and Quandt (1965) Test: This is a simple and intuitive test. One orders the observations according to Xi and omits c central observations. Next, two regressions are run on the two separated sets of observations with (n − c)/2 observations in each. The c omitted observations separate the low value X’s from the high value X’s, and if heteroskedasticity exists and is related to Xi , the estimates of σ 2 reported from the two regressions should be different. Hence, the test statistic is s22 /s21 where s21 and s22 are the Mean Square Error of the two regressions, respectively. Their ratio would be the same as that of the two residual sums of squares because the degrees of freedom of the two regressions are the same. This statistic is F distributed with ((n−c)/2)−K degrees of freedom in the numerator as well as the denominator. The only remaining question for performing this test is the magnitude of c. Obviously, the larger c is, the more central observations are being omitted and the more confident we feel that the two samples are distant from each other. The loss of c observations should lead to loss of power. However, separating the two samples should give us more confidence that the two variances are in fact the same if we do not reject homoskedasticity. This trade off in power was studied by Goldfeld and Quandt using Monte Carlo experiments. Their results recommend the use of c = 8 for n = 30 and c = 16 for n = 60. This is a popular test, but assumes that we know how to order the heteroskedasticity. In this case, using Xi . But what if there are more than one regressor on the right hand side? In that case one can order the observations using Yi . (3) Spearman’s Rank Correlation Test: This test ranks the Xi ’s and the absolute value of the OLS residuals, the ei ’s. Then it computes the difference between these rankings, i.e., di =

 rank(|ei |)− rank(Xi ). The Spearman-Correlation coefficient is r = 1 − 6 ni=1 d2i /(n3 − n) . Finally, test H0 ; the correlation coefficient between the rankings is zero, by computing t =


5.5

Heteroskedasticity

105

[r 2 (n − 2)/(1 − r 2 )]1/2 which is t-distributed with (n − 2) degrees of freedom. If this t-statistic has a large p-value we do not reject homoskedasticity. Otherwise, we reject homoskedasticity in favor of heteroskedasticity. (4) Harvey’s (1976) Multiplicative Heteroskedasticity Test: If heteroskedasticity is related to Xi , it looks like the Goldfeld and Quandt test or the Spearman rank correlation test would detect it, and the Glejser test would establish its form. In case the form of heteroskedasticity is of the multiplicative type, Harvey (1976) suggests the following test which rewrites (5.17) as loge2i = logσ 2 + δ 1 logZ1i + . . . + δr logZri + vi

(5.25)

where vi = log(e2i /σ 2i ). This disturbance term has an asymptotic distribution that is logχ21 . This random variable has mean −1.2704 and variance 4.9348. Therefore, Harvey suggests performing the regression in (5.25) and testing H0 ; δ 1 = δ 2 = . . . = δ r = 0 by computing the regression sum of squares divided by 4.9348. This statistic is distributed asymptotically as χ2r . This is also asymptotically equivalent to an F -test that tests for δ 1 = δ 2 = . . . = δr = 0 in the regression given in (5.25). See the F -test described in example 6 of Chapter 4. (5) Breusch and Pagan (1979) Test: If one knows that σ 2i = f (a + b1 Z1 + b2 Z2 + .. + br Zr ) but does not know the form of this function f , Breusch and Pagan (1979) suggest following n the 2 2 test for homoskedasticity, i.e., H0 ; b1 = b2 = . . . = br = 0. Compute σ  = i=1 ei /n, which σ 2 on the would be the MLE estimator of σ 2 under homoskedasticity. Run the regression of e2i / Z variables and a constant, and compute half the regression sum of squares. This statistic is distributed as χ2r . This is a more general test than the ones discussed earlier in that f does not have to be specified. (6) White’s (1980) Test: Another general test for homoskedasticity where nothing is known about the form of this heteroskedasticity is suggested by White (1980). This test is based on the difference between the variance of the OLS estimates under homoskedasticity and that under heteroskedasticity. For the case of a simple regression with a constant, White shows that  OLS ) given by (5.13) with the usual var(β  OLS ) = s2 / n x2 this test compares White’s var(β i=1 i under homoskedasticity. This test reduces to running the regression of e2i on a constant, Xi and Xi2 and computing nR2 . This statistic is distributed as χ22 under the null hypothesis of homoskedasticity. The degrees of freedom correspond to the number of regressors without the constant. If this statistic is not significant, then e2i is not related to Xi and Xi2 and we can not reject that the variance is constant. Note that if there is no constant in the regression, we run e2i on a constant and Xi2 only, i.e., Xi is no longer in this regression and the degree of freedom of the test is 1. In general, White’s test is based on running e2i on the cross-product of all the X’s in the regression being estimated, computing nR2 , and comparing it to the critical value of χ2r where r is the number of regressors in this last regression excluding the constant. For the case of two regressors, X2 and X3 and a constant, White’s test is again based on nR2 for the regression of e2i on a constant, X2 , X3 , X22 , X2 X3 , X32 . This statistic is distributed as χ25 . White’s test is standard using EViews. After running the regression, click on residuals tests then choose White. This software gives the user a choice between including or excluding the cross-product terms like X2 X3 from the regression. This may be useful when there are many regressors. A modified Breusch-Pagan test was suggested by Koenker (1981) and Koenker and Bassett (1982). This attempts to improve the power of the Breusch-Pagan test, and make it more robust


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Chapter 5: Violations of the Classical Assumptions

to the non-normality of the disturbances. This amounts to multiplying the Breusch-Pagan statistic (half the regression sum of squares) by 2 σ 4 , and dividing it by n nthe 2second sample 2 2 2 2 moment of the squared residuals, i.e., (e − σ  ) /n, where σ  = i=1 i i=1 ei /n. Waldman (1983) showed that if the Zi ’s in the Breusch-Pagan test are in fact the Xi ’s and their crossproducts, as in White’s test, then this modified Breusch-Pagan test is exactly the nR2 statistic proposed by White. White’s (1980) test for heteroskedasticity without specifying its form lead to further work on estimators that are more efficient than OLS while recognizing that the efficiency of GLS may not be achievable, see Cragg (1992). Adaptive estimators have been developed by Carroll (1982) and Robinson (1987). These estimators assume no particular form of heteroskedasticity but nevertheless have the same asymptotic distribution as GLS based on the true σ 2i . Many Monte Carlo experiments were performed to study the performance of these and other tests of homoskedasticity. One such extensive study is that of Ali and Giaccotto (1984). Six types of heteroskedasticity specifications were considered; (i) σ 2i = σ 2 (iv) σ 2i = σ 2 Xi2

(ii) σ 2i = σ 2 |Xi | (v) σ 2i = σ 2 [E(Yi )]2

(iii) σ 2i = σ 2 |E(Yi )| (vi) σ 2i = σ 2 for i ≤ n/2 and σ 2i = 2σ 2 for i > n/2

Six data sets were considered, the first three were stationary and the last three were nonstationary (Stationary versus non-stationary regressors, are discussed in Chapter 14). Five models were entertained, starting with a model with one regressor and no intercept and finishing with a model with an intercept and 5 variables. Four types of distributions were imposed on the disturbances. These were normal, t, Cauchy and log normal. The first three are symmetric, but the last one is skewed. Three sample sizes were considered, n = 10, 25, 40. Various correlations between the disturbances were also entertained. Among the tests considered were tests 1, 2, 5 and 6 discussed in this section. The results are too numerous to summarize, but some of the major findings are the following: (1) The power of these tests increased with sample size and trendy nature or the variability of the regressors. It also decreased with more regressors and for deviations from the normal distribution. The results were mostly erratic when the errors were autocorrelated. (2) There were ten distributionally robust tests using OLS residuals named TROB which were variants of Glejser’s, White’s and Bickel’s tests. The last one being a nonparametric test not considered in this chapter. These tests were robust to both long-tailed and skewed distributions. (3) None of these tests has any significant power to detect heteroskedasticity which deviates substantially from the true underlying heteroskedasticity. For example, none of these tests was powerful in detecting heteroskedasticity of the sixth kind, i.e., σ 2i = σ 2 for i ≤ n/2 and σ 2i = 2σ 2 for i > n/2. In fact, the maximum power was 9%. (4) Ali and Giaccotto (1984) recommend any of the TROB tests for practical use. They note that the similarity among these tests is the use of squared residuals rather than the absolute value of the residuals. In fact, they argue that tests of the same form that use absolute value rather than squared residuals are likely to be non-robust and lack power. Empirical Example: For the Cigarette Consumption Data given in Table 3.2, the OLS regression yields: logC = 4.30 − 1.34 logP + 0.17 logY (0.91) (0.32) (0.20)

¯ 2 = 0.27 R


5.5

Heteroskedasticity

107

0.4

Residuals

0.2

0.0

â&#x20AC;&#x201C;0.2

â&#x20AC;&#x201C;0.4 4.4

4.6

4.8

5.0

5.2

LNY Figure 5.1 Plots of Residuals versus Log Y

Suspecting heteroskedasticity, we plotted the residuals from this regression versus logY in Figure 5.1. This ďŹ gure shows the dispersion of the residuals to decrease with increasing logY . Next, we performed several tests for heteroskedasticity studied in this chapter. The ďŹ rst is Glejserâ&#x20AC;&#x2122;s (1969) test. We ran the following regressions: |ei | = 1.16 â&#x2C6;&#x2019; 0.22 logY (0.46) (0.10) |ei | = â&#x2C6;&#x2019;0.95 + 5.13 (logY )â&#x2C6;&#x2019;1 (0.47) (2.23) |ei | = â&#x2C6;&#x2019;2.00 + 4.65 (logY )â&#x2C6;&#x2019;0.5 (0.93) (2.04) |ei | = 2.21 â&#x2C6;&#x2019; 0.96 (logY )0.5 (0.93) (0.42) The t-statistics on the slope coeďŹ&#x192;cient in these regressions are â&#x2C6;&#x2019;2.24, 2.30, 2.29 and â&#x2C6;&#x2019;2.26, respectively. All are signiďŹ cant with p-values of 0.03, 0.026, 0.027 and 0.029, respectively, indicating the rejection of homoskedasticity. The second test is the Goldfeld and Quandt (1965) test. The observations are ordered according to logY and c = 12 central observations are omitted. Two regressions are run on the ďŹ rst and last 17 observations. The ďŹ rst regression yields s21 = 0.04881 and the second regression yields s22 = 0.01554. This is a test of equality of variances and it is based on the ratio of two Ď&#x2021;2 random variables with 17 â&#x2C6;&#x2019; 3 = 14 degrees of freedom. In fact, s21 /s22 = 0.04881/0.01554 = 3.141 â&#x2C6;ź F14,14 under H0 . This has a p-value of 0.02 and rejects H0 at the 5% level. The third test is the Spearman rank correlation test. First one obtains the rank(logYi ) and rank(|ei |) and  2 3 compute di = rank|ei |â&#x2C6;&#x2019; rank|logYi |. From these r = 1 â&#x2C6;&#x2019; 6 46 i=1 di /(n â&#x2C6;&#x2019; n) = â&#x2C6;&#x2019;0.282 and


108

Chapter 5: Violations of the Classical Assumptions

Table 5.1 White Heteroskedasticity Test F-statistic Obs*R-squared Test Equation: Dependent Variable: Method: Sample: Included observations: Variable C LNP LNPˆ2 LNP*LNY LNY LNYˆ2 R-squared Adjusted R-squared S.E. of regression Sum squared resid Log likelihood Durbin-Watson stat

4.127779 15.65644

Probability Probability

0.004073 0.007897

RESIDˆ2 Least Squares 1 46 46 Coefficient 18.22199 9.506059 1.281141 –2.078635 –7.893179 0.855726 0.340357 0.257902 0.029778 0.035469 99.58660 1.853360

Std. Error 5.374060 3.302570 0.656208 0.727523 2.329386 0.253048

t-Statistic 3.390730 2.878382 1.952340 –2.857139 –3.388523 3.381670

Mean dependent var S.D. dependent var Akaike info criterion Schwarz criterion F-statistic Prob (F-statistic)

Prob. 0.0016 0.0064 0.0579 0.0068 0.0016 0.0016 0.024968 0.034567 –4.068982 –3.830464 4.127779 0.004073

t = [r 2 (n − 2)/(1 − r 2 )]1/2 = 1.948. This is distributed as a t with 44 degrees of freedom. This t-statistic has a p-value of 0.058. The fourth test is Harvey’s (1976) multiplicative heteroskedasticity test which is based upon regressing log e2i on log(log Yi ) log e2i = 24.85 − 19.08 log(log Y ) (17.25) (11.03) Harvey’s (1976) statistic divides the regression sum of squares which is 14.360 by 4.9348. This yields 2.91 which is asymptotically distributed as χ21 under the null. This has a p-value of 0.088 and does not reject the null of homoskedasticity at the 5% significance level. The fifth test is the Breusch and Pagan (1979) test which is based on the regression of e2i / σ2 46 2 2 (where σ  = i=1 ei /46 = 0.024968) on log Yi . The test-statistic is half the regression sum of squares = (10.971 ÷ 2) = 5.485. This is distributed as χ21 under the null hypothesis. This has a p-value of 0.019 and rejects the null of homoskedasticity. Finally, White’s (1980) test for heteroskedasticity is performed which is based on the regression of e2i on logP , logY , (logP )2 , (logY )2 , (logP )(logY ) and a constant. This is shown in Table 5.1 using EViews. The test-statistic is nR2 = (46)(0.3404) = 15.66 which is distributed as χ25 . This has a p-value of 0.008 and rejects the null of homoskedasticity. Except for Harvey’s test, all the tests performed indicate the presence of heteroskedasticity. This is true despite the fact that the data are in logs, and both consumption and income are expressed in per capita terms.


5.6

Autocorrelation

109

Table 5.2 White Heteroskedasticity-Consistent Standard Errors Dependent Variable: Method: Sample: Included observations:

LNC Least Squares 1 46 46

White Heteroskedasticity-Consistent Standard Errors & Covariance Variable C LNP LNY

Coefficient 4.299662 –1.338335 0.172386

R-squared Adjusted R-squared S.E. of regression Sum squared resid Log likelihood Durbin-Watson stat

0.303714 0.271328 0.163433 1.148545 19.60218 2.315716

Std. Error 1.095226 0.343368 0.236610

t-Statistic 3.925821 –3.897671 0.728565

Mean dependent var S.D. dependent var Akaike info criterion Schwarz criterion F-statistic Prob (F-statistic)

Prob. 0.0003 0.0003 0.4702 4.847844 0.191458 –0.721834 –0.602575 9.378101 0.000417

White’s heteroskedasticity-consistent estimates of the variances are as follows: logC = 4.30 − 1.34 logP + 0.17 logY (1.10) (0.34) (0.24) These are given in Table 5.2 using EViews. Note that in this case all of the heteroskedasticityconsistent standard errors are larger than those reported using a standard OLS package, but this is not necessarily true for other data sets. In section 5.4, we described the Jarque and Bera (1987) test for normality of the disturbances. For this cigarette consumption regression, Figure 5.2 gives the histogram of the residuals along with descriptive statistics of these residuals including their mean, median, skewness and kurtosis. This is done using EViews. The measure of skewness S is estimated to be −0.184 and the measure of kurtosis κ is estimated to be 2.875 yielding a Jarque-Bera statistic of

(2.875 − 3)2 (−0.184)2 = 0.29. + JB = 46 6 24 This is distributed as χ22 under the null hypothesis of normality and has a p-value of 0.865. Hence we do not reject that the distribution of the disturbances is symmetric and has a kurtosis of 3.

5.6

Autocorrelation

Violation of assumption 3 means that the disturbances are correlated, i.e., E(ui uj ) = σ ij = 0, for i = j, and i, j = 1, 2, . . . , n. Since ui has zero mean, E(ui uj ) = cov(ui , uj ) and this is denoted by σ ij . This correlation is more likely to occur in time-series than in cross-section studies. Consider estimating the consumption function of a random sample of households. An unexpected event, like a visit of family members will increase the consumption of this household. However, this


110

Chapter 5: Violations of the Classical Assumptions

10

Series: Residuals Sample 1 46 Observations 46

8

Mean Median Maximum Minimum Std. Dev. Skewness Kurtosis

6

4

–9.90E-16 0.007568 0.328677 –0.418675 0.159760 –0.183945 2.875020

2

Jarque-Bera Probability

0.289346 0.865305

0 –0.4

–0.3

–0.2

–0.1

0.0

0.1

0.2

0.3

Figure 5.2 Normality Test (Jarque-Bera)

positive disturbance need not be correlated to the disturbances affecting consumption of other randomly drawn households. However, if we were estimating this consumption function using aggregate time-series data for the U.S., then it is very likely that a recession year affecting consumption negatively this year may have a carry over effect to the next few years. A shock to the economy like an oil embargo in 1973 is likely to affect the economy for several years. A labor strike this year may affect production for the next few years. Therefore, we will switch the i and j subscripts to t and s denoting time-series observations t, s = 1, 2, . . . , T and the sample size will be denoted by T rather than n. This covariance term is symmetric, so that σ 12 = E(u1 u2 ) = E(u2 u1 ) = σ 21 . Hence, only T (T − 1)/2 distinct σ ts ’s have to be estimated. For example, if T = 3, then σ 12 , σ 13 and σ 23 are the distinct covariance terms. However, it is hopeless to estimate T (T − 1)/2 covariances (σ ts ) with only T observations. Therefore, more structure on these σ ts ’s need to be imposed. A popular assumption is that the ut ’s follow a first-order autoregressive process denoted by AR(1): ut = ρut−1 + ǫt

t = 1, 2, . . . , T

(5.26)

where ǫt is IID(0, σ 2ǫ ). It is autoregressive because ut is related to its lagged value ut−1 . One can also write (5.26), for period t − 1, as ut−1 = ρut−2 + ǫt−1

(5.27)

and substitute (5.27) in (5.26) to get ut = ρ2 ut−2 + ρǫt−1 + ǫt

(5.28)

Note that the power of ρ and the subscript of u or ǫ always sum to t. By continuous substitution of this form, one ultimately gets ut = ρt u0 + ρt−1 ǫ1 + .. + ρǫt−1 + ǫt

(5.29)


5.6

Autocorrelation

111

This means that ut is a function of current and past values of ǫt and u0 where u0 is the initial value of ut . If u0 has zero mean, then ut has zero mean. This follows from (5.29) by taking expectations. Also, from (5.26) var(ut ) = ρ2 var(ut−1 ) + var(ǫt ) + 2ρcov(ut−1 , ǫt )

(5.30)

Using (5.29), ut−1 is a function of ǫt−1 , past values of ǫt−1 and u0 . Since u0 is independent of the ǫ’s, and the ǫ’s are themselves not serially correlated, then ut−1 is independent of ǫt . This means that cov(ut−1 , ǫt ) = 0. Furthermore, for ut to be homoskedastic, var(ut ) = var(ut−1 ) = σ 2u , and (5.30) reduces to σ 2u = ρ2 σ 2u + σ 2ǫ , which when solved for σ 2u gives: σ 2u = σ 2ǫ /(1 − ρ2 )

(5.31)

Hence, u0 ∼ (0, σ 2ǫ /(1 − ρ2 )) for the u’s to have zero mean and homoskedastic disturbances. Multiplying (5.26) by ut−1 and taking expected values, one gets E(ut ut−1 ) = ρE(u2t−1 ) + E(ut−1 ǫt ) = ρσ 2u

(5.32)

since E(u2t−1 ) = σ 2u and E(ut−1 ǫt ) = 0. Therefore, cov(ut , ut−1 ) = ρσ 2u , and the correlation coef ficient between ut and ut−1 is correl(ut , ut−1 ) = cov(ut , ut−1 )/ var(ut )var(ut−1 ) = ρσ 2u /σ 2u = ρ. Since ρ is a correlation coefficient, this means that −1 ≤ ρ ≤ 1. In general, one can show that cov(ut , us ) = ρ|t−s| σ 2u

t, s = 1, 2, . . . , T

(5.33)

see problem 6. This means that the correlation between ut and ut−r is ρr , which is a fraction raised to an integer power, i.e., the correlation is decaying between the disturbances the further apart they are. This is reasonable in economics and may be the reason why this autoregressive form (5.26) is so popular. One should note that this is not the only form that would correlate the disturbances across time. In Chapter 14, we will consider other forms like the Moving Average (MA) process, and higher order Autoregressive Moving Average (ARMA) processes, but these are beyond the scope of this chapter. Consequences for OLS How is the OLS estimator affected by the violation of the no autocorrelation assumption among the disturbances? The OLS estimator is still unbiased and consistent since these properties rely on assumptions 1 and 4 and have nothing to do with assumption 3. For the simple linear  regression, using (5.2), the variance of β OLS is now  T T T  (5.34) var(β OLS ) = var( t=1 wt ut ) = t=1 s=1 wt ws cov(ut , us )  2 T 2 |t−s| 2 = σ u / t=1 xt + wt ws ρ σu t=s

where cov(ut , us ) = ρ|t−s| σ 2u as explained in (5.33). Note that the first term in (5.34) is the  usual variance of β OLS under the classical case. The second term in (5.34) arises because of the correlation between the ut ’s. Hence, the variance of OLS computed from a regression package,   i.e., s2 / Tt=1 x2t is a wrong estimate of the variance of β OLS for two reasons. First, it is using T 2 2 the wrong formula for the variance, i.e., σ u / t=1 xt rather than (5.34). The latter depends on ρ through the extra term in (5.34). Second, one can show, see problem 7, that E(s2 ) = σ 2u and will


112

Chapter 5: Violations of the Classical Assumptions

 involve ρ as well as σ 2u . Hence, s2 is not unbiased for σ 2u and s2 / Tt=1 x2t is a biased estimate of  var(β In fact, if ρ OLS ). The direction and magnitude of this bias depends on ρ and the regressor.  is positive, and the xt ’s are themselves positively autocorrelated, then s2 / Tt=1 x2t understates  the true variance of β OLS . This means that the confidence interval for β is tighter than it should be and the t-statistic for H0 ; β = 0 is overblown, see problem 8. As in the heteroskedastic  case, but for completely different reasons, any inference based on var(β OLS ) reported from the standard regression packages will be misleading if the ut ’s are serially correlated. Newey and West (1987) suggested a simple heteroskedasticity and autocorrelation-consistent covariance matrix for the OLS estimator without specifying the functional form of the serial correlation. The basic idea extends White’s (1980) replacement of heteroskedastic variances with squared OLS residuals e2t by additionally including products of least squares residuals et et−s for s = 0, ±1, . . . , ±p where p is the maximum order of serial correlation we are willing to assume. The consistency of this procedure relies on p being very small relative to the number of observations T . This is consistent with popular serial correlation specifications considered in this chapter where the autocorrelation dies out quickly as j increases. Newey and West (1987) allow the higher order covariance terms to receive diminishing weights. This NeweyWest option for the least squares estimator is available using EViews. Andrews (1991) warns about the unreliability of such standard error corrections in some circumstances. Wooldridge (1991) shows that it is possible to construct serially correlated robust F -statistics for testing joint hypotheses as considered in Chapter 4. However, these are beyond the scope of this book. Is OLS still BLUE? In order to determine the BLU estimator in this case, we lag the regression equation once, multiply it by ρ, and subtract it from the original regression equation, we get Yt − ρYt−1 = α(1 − ρ) + β(Xt − ρXt−1 ) + ǫt

t = 2, 3, . . . , T

(5.35)

This transformation, known as the Cochrane-Orcutt (1949) transformation, reduces the disturbances to classical errors. Therefore, OLS on the resulting regression renders the estimates t = Xt − ρXt−1 , for t = 2, 3, . . . , T . Note BLU, i.e., run Yt = Yt − ρYt−1 on a constant and X that we have lost one observation by lagging, and the resulting estimators are BLUE only for linear combinations of (T − 1) observations in Y .1 Prais and Winsten (1954) derive the BLU estimators for linear combinations of T observations in Y . This entails recapturing the  initial observation as follows: (i) Multiply the first observation of the regression equation by 1 − ρ2 ;     1 − ρ2 Y1 = α 1 − ρ2 + β 1 − ρ2 X1 + 1 − ρ2 u1 (ii) add this transformed initial observation to the Cochrane-Orcutt transformed observations for t = 2, . . . , T and run the regression on the T observations rather than the (T −1) observations. See Chapter 9, for a formal proof of this result. Note that  Y1 = 1 − ρ2 Y1

and

Yt = Yt − ρYt−1 for t = 2, . . . , T  1 = 1 − ρ2 X1 and X t = Xt − ρXt−1 for t = 2, . . . , T . The constant variable Ct = 1 Similarly, X  t = (1 − ρ) t which takes the values C 1 = 1 − ρ2 and C for t = 1, . . . , T is now a new variable C  t and X t for t = 2, . . . , T . Hence, the Prais-Winsten procedure is the regression of Yt on C


5.6

Autocorrelation

113

without a constant. It is obvious that the resulting BLU estimators will involve ρ and are therefore, different from the usual OLS estimators except in the case where ρ = 0. Hence, OLS is no longer BLUE. Furthermore, we need to know ρ in order to obtain the BLU estimators. In applied work, ρ is not known and has to be estimated, in which case the Prais-Winsten regression is no longer BLUE since it is based on an estimate of ρ rather than the true ρ itself. However, as long as  ρ is a consistent estimate for ρ then this is a sufficient condition for the corresponding estimates of α and β in the next step to be asymptotically efficient, see Chapter 9. We now turn to various methods of estimating ρ. (1) The Cochrane-Orcutt (1949) Method: This method starts with an initial estimate of ρ, the most convenient is 0, and minimizes the residual sum of squares in (5.35). This gives us  the OLS estimates of α and β. Then we substitute α  OLS and β OLS in (5.35) and we get et = ρet−1 + ǫt

t = 2, . . . , T

(5.36)

where et denotes the OLS residual. An estimate of ρ can be obtained by minimizing the residual sum of squares in (5.36) or running the  regression of et on et−1 without a constant. The resulting estimate of ρ is  ρco = Tt=2 et et−1 / Tt=2 e2t−1 where both summations run over t = 2, 3, . . . , T . The second step of the Cochrane-Orcutt procedure (2SCO) is to perform the regression in (5.35) with  ρco instead of ρ. One can iterate this procedure (ITCO) by computing new residuals based on the new estimates of α and β and hence a new estimate of ρ from (5.36), and so on, until convergence. Both the 2SCO and the ITCO are asymptotically efficient, the argument for iterating must be justified in terms of small sample gains. (2) The Hilderth-Lu (1960) Search Procedure: ρ is between −1 and 1. Therefore, this procedure searches over this range, i.e., using values of ρ say between −0.9 and 0.9 in intervals of 0.1. For each ρ, one computes the regression in (5.35) and reports the residual sum of squares corresponding to that ρ. The minimum residual sum of squares gives us our choice of ρ and the corresponding regression gives us the estimates of α, β and σ 2 . One can refine this procedure around the best ρ found in the first stage of the search. For example, suppose that ρ = 0.6 gave the minimum residual sum of squares, one can search next between 0.51 and 0.69 in intervals of 0.01. This search procedure guards against a local minimum. Since the likelihood in this case contains ρ as well as σ 2 and α and β, this search procedure can be modified to maximize the likelihood rather than minimize the residual sum of squares, since the two criteria are no longer equivalent. The maximum value of the likelihood will give our choice of ρ and the corresponding estimates of α, β and σ 2 . (3) Durbin’s (1960) Method: One can rearrange (5.35) by moving Yt−1 to the right hand side, i.e., Yt = ρYt−1 + α(1 − ρ) + βXt − ρβXt−1 + ǫt

(5.37)

and running OLS on (5.37). The error in (5.37) is classical, and the presence of Yt−1 on the right hand side reminds us of the contemporaneously uncorrelated case discussed under the violation of assumption 4. For that violation, we have shown that unbiasedness is lost, but not consistency. Hence, the estimate of ρ as a coefficient of Yt−1 is biased but consistent. This is the Durbin estimate of ρ, call it  ρD . Next, the second step of the Cochrane-Orcutt procedure is performed using this estimate of ρ.


114

Chapter 5: Violations of the Classical Assumptions

(4) Beach-MacKinnon (1978) Maximum Likelihood Procedure: Beach and MacKinnon (1978) derived a cubic equation in ρ which maximizes the likelihood function concentrated with respect to α, β, and σ 2 . With this estimate of ρ, denoted by  ρBM , one performs the PraisWinsten procedure in the next step. Correcting for serial correlation is not without its critics. Mizon (1995) argues this point forcefully in his article entitled “A simple message for autocorrelation correctors: Don’t.” The main point being that serial correlation is a symptom of dynamic misspecification which is better represented using a general unrestricted dynamic specification. Monte Carlo Results Rao and Griliches (1969) performed a Monte Carlo study using an autoregressive Xt , and various values of ρ. They found that OLS is still a viable estimator as long as |ρ| < 0.3, but if |ρ| > 0.3, then it pays to perform procedures that correct for serial correlation based on an estimator of ρ. Their recommendation was to compute a Durbin’s estimate of ρ in the first step and to do the Prais-Winsten procedure in the second step. Maeshiro (1976, 1979) found that if the Xt series is trended, which is usual with economic data, then OLS outperforms 2SCO, but not the two-step Prais-Winsten (2SPW) procedure that recaptures the initial observation. These results were confirmed by Park and Mitchell (1980) who performed an extensive Monte Carlo using trended and untrended Xt ’s. Their basic findings include the following: (i) For trended Xt ’s, OLS beats 2SCO, ITCO and even a Cochrane-Orcutt procedure that is based on the true ρ. However, OLS was beaten by 2SPW, iterative Prais-Winsten (ITPW), and BeachMacKinnon (BM). Their conclusion is that one should not use regressions based on (T − 1) observations as in Cochrane and Orcutt. (ii) Their results find that the ITPW procedure is the recommended estimator beating 2SPW and BM for high values of true ρ, for both trended as well as nontrended Xt ’s. (iii) Test of hypotheses regarding the regression coefficients performed miserably for all estimators based on an estimator of ρ. The results indicated less bias in standard error estimation for ITPW, BM and 2SPW than OLS. However, the tests based on these standard errors still led to a high probability of type I error for all estimation procedures. Testing for Autocorrelation So far, we have studied the properties of OLS under the violation of assumption 3. We have derived asymptotically efficient estimators of the coefficients based on consistent estimators of ρ and studied their small sample properties using Monte Carlo experiments. Next, we focus on the problem of detecting this autocorrelation between the disturbances. A popular diagnostic for detecting such autocorrelation is the Durbin and Watson (1951) statistic2 d=

T

t=2 (et

− et−1 )2 /

T

2 t=1 et

(5.38)

If this was based on the true ut ’s and T was very large then d can be shown to tend in the limit as T gets large to 2(1 − ρ), see problem 9. This means that if ρ → 0, then d → 2; if ρ → 1, then d → 0 and if ρ → −1, then d → 4. Therefore, a test for H0 ; ρ = 0, can be based on whether d is close to 2 or not. Unfortunately, the critical values of d depend upon the Xt ’s, and these vary from one data set to another. To get around this, Durbin and Watson established upper (dU ) and lower (dL ) bounds for this critical value. Figure 5.3 shows these bounds.


5.6

Autocorrelation

115

f(d)

Reject H 0

Inconclusive

Inconclusive

Do not reject H 0

Reject H 0

d 0

d1

du

2

4– d u

4– d1

4

Figure 5.3 Durbin-Watson Critical Values

It is obvious that if the observed d is less than dL , or larger than 4 − dL , we reject H0 . If the observed d is between dU and 4 − dU , then we do not reject H0 . If d lies in any of the two indeterminant regions, then one should compute the exact critical values depending on Xt . Most regression packages report the Durbin-Watson statistic. SHAZAM gives the exact p-value for this d-statistic. If one is interested in a single sided test, say H0 ; ρ = 0 versus H1 ; ρ > 0 then one would reject H0 if d < dL , and not reject H0 if d > dU . If dL < d < dU , then the test is inconclusive. Similarly for testing H0 ; ρ = 0 versus H1 ; ρ < 0, one computes (4 − d) and follow the steps for testing against positive autocorrelation. Durbin and Watson tables for dL and dU covered samples sizes from 15 to 100 and a maximum of 5 regressors. Savin and White (1977) extended these tables for 6 ≤ T ≤ 200 and up to 10 regressors. The Durbin-Watson statistic has several limitations. We discussed the inconclusive region and the computation of exact critical values. The Durbin-Watson statistic is appropriate when there is a constant in the regression. In case there is no constant in the regression, see Farebrother (1980). Also, the Durbin-Watson statistic is inappropriate when there are lagged values of the dependent variable among the regressors. We now turn to an alternative test for serial correlation that does not have these limitations and that is also easy to apply. This test was derived by Breusch (1978) and Godfrey (1978) and is known as the Breusch-Godfrey test for zero first-order serial correlation. This is a Lagrange Multiplier test that amounts to running the regression of the OLS residuals et on et−1 and the original regressors in the model. The test statistic is T R2 . Its distribution under the null is χ21 . In this case, the regressors are a constant and Xt , and the test checks whether the coefficient of et−1 is significant. The beauty of this test is that (i) it is the same test for first-order serial correlation, whether the disturbances are Moving Average of order one MA(1) or AR(1). (ii) This test is easily generalizable to higher autoregressive or Moving Average schemes. For second-order serial correlation, like MA(2) or AR(2) one includes two lags of the residuals on the right hand side; i.e., both et−1 and et−2 . (iii) This test is still valid even when lagged values of the dependent variable are present among the regressors, see Chapter 6. The Breusch and Godfrey test is standard using EViews and it prompts the user


116

Chapter 5: Violations of the Classical Assumptions

with a choice of the number of lags of the residuals to include among the regressors to test for serial correlation. You click on residuals, then tests and choose Breusch-Godfrey. Next, you input the number of lagged residuals you want to include. In conclusion, we focus on first differencing the data as a possible solution for getting rid of serial correlation in the errors. Some economic behavioral equations have variables in first difference form, but other equations are first differenced for estimation purposes. In the latter case, if the original disturbances were not autocorrelated, (or even correlated, with ρ = 1), then the transformed disturbances are serially correlated. After all, first differencing the disturbances is equivalent to setting ρ = 1 in ut − ρut−1 , and this new disturbance u∗t = ut − ut−1 has ut−1 in common with u∗t−1 = ut−1 − ut−2 , making E(u∗t u∗t−1 ) = −E(u2t−1 ) = −σ 2u . However, one could argue that if ρ is large and positive, first differencing the data may not be a bad solution. Rao and Miller (1971) calculated the variance of the BLU estimator correcting for serial correlation, for various guesses of ρ. They assume a true ρ of 0.2, and an autoregressive Xt Xt = λXt−1 + wt

with λ = 0, 0.4, 0.8.

(5.39)

They find that OLS (or a guess of ρ = 0), performs better than first differencing the data, and is pretty close in terms of efficiency to the true BLU estimator for trended Xt (λ = 0.8). However, the performance of OLS deteriorates as λ declines to 0.4 and 0, with respect to the true BLU estimator. This supports the Monte Carlo finding by Rao and Griliches that for |ρ| < 0.3, OLS performs reasonably well relative to estimators that correct for serial correlation. However, the first-difference estimator, i.e., a guess of ρ = 1, performs badly for trended Xt (λ = 0.8) giving the worst efficiency when compared to any other guess of ρ. Only when the Xt ’s are less trended (λ = 0.4) or random (λ = 0), does the efficiency of the first-difference estimator improve. However, even for those cases one can do better by guessing ρ. For example, for λ = 0, one can always do better than first differencing by guessing any positive ρ less than 1. Similarly, for true ρ = 0.6, a higher degree of serial correlation, Rao and Miller (1971) show that the performance of OLS deteriorates, while that of the first difference improves. However, one can still do better than first differencing by guessing in the interval (0.4, 0.9). This gain in efficiency increases with trended Xt ’s. Empirical Example: Table 5.3 gives the U.S. Real Personal Consumption Expenditures (C) and Real Disposable Personal Income (Y ) from the Economic Report of the President over the period 1950-1993. This data set is available as CONSUMP.DAT on the Springer web site. The OLS regression yields: Ct = −65.80 + 0.916 Yt + residuals (90.99) (0.009) Figure 5.4 plots the actual, fitted and residuals using EViews. This shows positive serial correlation with a string of positive residuals followed by a string of negative residuals followed by positive residuals. The Durbin-Watson statistic is d = 0.461 which is much smaller than the lower bound dL = 1.468 for T = 44 and one regressor. Therefore, we reject the null hypothesis of H0 ; ρ = 0 at the 5% significance level. Regressing OLS residuals on their lagged values yields et = 0.792 et−1 + residuals (0.106)


5.6

Autocorrelation

117

Table 5.3 U.S. Consumption Data, 1950–1993 C = Real Personal Consumption Expenditures (in 1987 dollars) Y = Real Disposable Personal Income (in 1987 dollars) YEAR

Y

C

YEAR

Y

C

1950 1951 1952 1953 1954 1955 1956 1957 1958 1959 1960 1961 1962 1963 1964 1965 1966 1967 1968 1969 1970 1971

6284 6390 6476 6640 6628 6879 7080 7114 7113 7256 7264 7382 7583 7718 8140 8508 8822 9114 9399 9606 9875 10111

5820 5843 5917 6054 6099 6325 6440 6465 6449 6658 6698 6740 6931 7089 7384 7703 8005 8163 8506 8737 8842 9022

1972 1973 1974 1975 1976 1977 1978 1979 1980 1981 1982 1983 1984 1985 1986 1987 1988 1989 1990 1991 1992 1993

10414 11013 10832 10906 11192 11406 11851 12039 12005 12156 12146 12349 13029 13258 13552 13545 13890 14005 14101 14003 14279 14341

9425 9752 9602 9711 10121 10425 10744 10876 10746 10770 10782 11179 11617 12015 12336 12568 12903 13029 13093 12899 13110 13391

Source: Economic Report of the President

The second step of the Cochrane-Orcutt (1949) procedure based on  ρ = 0.792 yields the following regression: (Ct − 0.792Ct−1 ) = 168.49 + 0.926 (Yt − 0.792Yt−1 ) + residuals (301.7) (0.027)

The two-step Prais-Winsten (1954) procedure yields  ρP W = 0.707 with standard error (0.110) and the regression estimates are given by Ct = −52.63 + 0.917 Yt + residuals (181.9) (0.017)

Iterative Prais-Winsten yields Ct = −48.40 + 0.916 Yt + residuals (191.1) (0.018) The Maximum-Likelihood procedure yields  ρM LE = 0.788 with standard error (0.11) and the resulting estimates are given by Ct = −24.82 + 0.915 Yt + residuals (233.5) (0.022)


118

Chapter 5: Violations of the Classical Assumptions

14000 12000 10000 400

8000

200

6000 4000

0 –200 –400 50

55

60

65

70

Residual

75

80

Actual

85

90

Fitted

Figure 5.4 Consumption and Disposable Income

Table 5.4 Breusch-Godfrey LM Test F-statistic Obs*R-squared Test Equation: Dependent Variable: Method:

53.45697 24.90136

Probability Probability

0.00000 0.000001

RESID Least Squares

Presample missing value lagged residuals set to zero Variable

Coefficient

Std. Error

t-Statistic

Prob.

C Y RESID(–1)

–31.12035 0.003641 0.800205

60.82348 0.005788 0.109446

–0.511650 0.629022 7.311428

0.6116 0.5328 0.0000

R-squared Adjusted R-squared S.E. of regression Sum squared resid Log likelihood Durbin-Watson stat

0.565940 0.544766 102.4243 430120.2 –264.5612 1.973552

Mean dependent var S.D. dependent var Akaike info criterion Schwarz criterion F-statistic Prob (F-statistic)

9.83E-13 151.8049 12.16187 12.28352 26.72849 0.000000


Notes

119

Table 5.5 Newey-West Standard Errors Dependent Variable: Method: Sample: Included observations:

CONSUM Least Squares 1950 1993 44

Newey-West HAC Standard Errors & Covariance (lag truncation=3) Variable C Y

Coefficient –65.79582 0.915623

Std. Error 133.3454 0.015458

t-Statistic –0.493424 59.23190

R-squared Adjusted R-squared S.E. of regression Sum squared resid Log likelihood Durbin-Watson stat

0.996267 0.996178 153.6015 990923.1 –282.9218 0.460778

Mean dependent var S.D. dependent var Akaike info criterion Schwarz criterion F-statistic Prob (F-statistic)

Prob. 0.6243 0.0000 9250.545 2484.624 12.95099 13.03209 11209.21 0.000000

Durbin’s (1960) Method yields the following regression Ct = 0.80 Ct−1 − 40.79 + 0.72 Yt − 0.53 Yt−1 + residuals (0.10) (59.8) (0.09) (0.13) Therefore, Durbin’s estimate of ρ is given by  ρD = 0.80. The Breusch (1978) and Godfrey (1978) regression that tests for first-order serial correlation is given in Table 5.4. This yields et = −31.12 + 0.004 Yt + 0.800 et−1 + residuals (60.82) (0.006) (0.109) The test statistic is T R2 which yields 43 × (0.565) = 24.9. This is distributed as χ21 under H0 ; ρ = 0. This rejects the null hypothesis of no serial correlation with a p-value of 0.000001 shown in Table 5.4. The Newey-West heteroskedasticity and autocorrelation-consistent standard errors for least squares with a three-year lag truncation are given by Ct = −65.80 + 0.916 Yt + residuals (133.3) (0.015) This is given in Table 5.5 using EViews. Note that both standard errors are now larger than those reported by least squares. But once again, this is not necessarily the case for other data sets.

Notes 1. A computational warning is in order when one is applying the Cochrane-Orcutt transformation to cross-section data. Time-series data has a natural ordering which is generally lacking in crosssection data. Therefore, one should be careful in applying the Cochrane-Orcutt transformation to cross-section data since it is not invariant to the ordering of the observations.


120

Chapter 5: Violations of the Classical Assumptions

2. Another test for serial correlation can be obtained as a by-product of maximum likelihood estimation. The maximum likelihood estimator of ρ has a normal limiting distribution with mean ρ and variance (1 − ρ2 )/T . Hence, one can compute  ρMLE /[(1 −  ρ2MLE )/T ]1/2 and compare it to critical values from the normal distribution.

Problems 1. For the simple linear regression with heteroskedasticity, i.e., E(u2i ) = σ 2i , show that E(s2 ) is a function of the σ 2i ’s? 2. For the simple linear regression with heteroskedasticity of the form E(u2i ) = σ 2i = bx2i where b > 0, n  show that E(s2 / i=1 x2i ) understates the variance of β OLS which is n  n 2 2 2 2 i=1 xi ) . i=1 xi σ i /(

3. Weighted Least Squares. This is based on Kmenta (1986).

(a) Solve the two equations in (5.11) and show that the solution is given by (5.12). (b) Show that n 2 i=1 (1/σ i )  n  n n 2 2 2 [ i=1 Xi /σ i ][ i=1 (1/σi )] − [ i=1 (Xi /σ 2i )]2 n ∗ w  i=1 i  = n ( i=1 wi∗ Xi2 )( ni=1 wi∗ ) − ( ni=1 wi∗ Xi )2 1 = n ∗ ¯∗ 2 i=1 wi (Xi − X )  ¯ ∗ = n w∗ Xi / n w∗ . where wi∗ = (1/σ 2i ) and X i=1 i i=1 i  var(β)

=

4. Relative Efficiency of OLS Under Heteroskedasticity. Consider the simple linear regression with heteroskedasticity of the form σ 2i = σ 2 Xiδ where Xi = 1, 2, . . . , 10.  (a) Compute var(β OLS ) for δ = 0.5, 1, 1.5 and 2.  (b) Compute var(β BLUE ) for δ = 0.5, 1, 1.5 and 2.    (c) Compute the efficiency of β OLS = var(β BLUE )/var(β OLS ) for δ = 0.5, 1, 1.5 and 2. What happens to this efficiency measure as δ increases?

5. Consider the simple regression with only a constant yi = α + ui for i = 1, 2, . . . , n; where the ui ’s are independent with mean zero and var(ui ) = σ 21 for i = 1, 2, . . . , n1 ; and var(ui ) = σ 22 for i = n1 + 1, . . . , n1 + n2 with n = n1 + n2 . (a) Derive the OLS estimator of α along with its mean and variance. (b) Derive the GLS estimator of α along with its mean and variance. (c) Obtain the relative efficiency of OLS with respect to GLS. Compute their relative efficiency for various values of σ 22 /σ 21 = 0.2, 0.4, 0.6, 0.8, 1, 1.25, 1.33, 2.5, 5; and n1 /n = 0.2, 0.3, 0.4, . . . , 0.8. Plot this relative efficiency. (d) Assume that ui is N (0, σ 21 ) for i = 1, 2, . . . , n1 ; and N (0, σ 21 ) for i = n1 + 1, . . . , n1 + n2 ; with ui ’s being independent. What is the maximum likelihood estimator of α, σ21 and σ 22 ? (e) Derive the LR test for testing H0 ; σ 21 = σ 22 in part (d).


Problems

121

6. Show that for an AR(1) model given in (5.26), E(ut us ) = ρ|t−s| σ 2u for t, s = 1, 2, . . . , T . 7. Relative Efficiency of OLS Under the AR(1) Model. This problem is based on Johnston (1984, pp. 310-312). For the simple regression without a constant yt = βxt + ut with ut = ρut−1 + ǫt and ǫt ∼ IID(0, σ 2ǫ ) (a) Show that  var(β OLS )

σ 2u T

=

1 + 2ρ

2 t=1 xt

T −1

+ . . . + 2ρ

T −1

t=1 xt xt+1 T 2 t=1 xt

x1 xT T 2 t=1 xt

+ 2ρ2

!

T −2

t=1 xt xt+2 T 2 t=1 xt

 and that the Prais-Winsten estimator β P W has variance   σ 2u 1 − ρ2  var(β P W ) = T T −1 T 2 2 1 + ρ2 − 2ρ t=1 xt xt+1 / t=1 xt t=1 xt

These expressions are easier to prove using matrix algebra, see Chapter 9.

(b) Let xt itself follow an AR(1) scheme with parameter λ, i.e., xt = λxt−1 + vt , and let T → ∞. Show that  var(β 1 − ρ2 PW )  = lim asy eff(β OLS ) = 2  T →∞ var(β (1 + ρ − 2ρλ)(1 + 2ρλ + 2ρ2 λ2 + . . .) OLS ) =

(1 − ρ2 )(1 − ρλ) (1 + ρ2 − 2ρλ)(1 + ρλ)

 (c) Tabulate this asy eff(β OLS ) for various values of ρ and λ where ρ varies between −0.9 to +0.9 in increments of 0.1, while λ varies between 0 and 0.9 in increments of 0.1. What do you conclude? How serious is the loss in efficiency in using OLS rather than the PW procedure? T  (d) Ignoring this autocorrelation one would compute σ 2u / t=1 x2t as the var(β OLS ). The difference between this wrong formula and that derived in part (a) gives us the bias in estimating  the variance of β OLS . Show that as T → ∞, this asymptotic proportionate bias is given by −2ρλ/(1 + ρλ). Tabulate this asymptotic bias for various values of ρ and λ as in part (c). What do you conclude? How serious is the asymptotic bias of using the wrong variances for  β OLS when the disturbances are first-order autocorrelated? (e) Show that

2

E(s ) =

σ 2u



T−

1 + 2ρ T −1

+ . . . + 2ρ

T −1

t=1 xt xt+1 T 2 t=1 xt

x1 xT T 2 t=1 xt

!

+ 2ρ2

/(T − 1)

T −2

t=1 xt xt+2 T 2 t=1 xt

Conclude that if ρ = 0, then E(s2 ) = σ 2u . If xt follows an AR(1) scheme with parameter λ, then for a large T , we get   1 + ρλ E(s2 ) = σ 2u T − /(T − 1) 1 − ρλ Compute this E(s2 ) for T = 101 and various values of ρ and λ as in part (c). What do you conclude? How serious is the bias in using s2 as an unbiased estimator for σ 2u ?


122

Chapter 5: Violations of the Classical Assumptions

8. For the AR(1) model given in (5.26), show that if ρ > 0 and the xt ’s are positively autocorrelated   that E(s2 / x2t ) understates the var(β OLS ) given in (5.34). 9. Show that for the AR(1) model, the Durbin-Watson statistic has plimd → 2(1 − ρ).

10. Regressions with Non-Zero Mean Disturbances. Consider the simple regression with a constant Yi = α + βXi + ui

i = 1, 2, . . . , n

where α and β are scalars and ui is independent of the Xi ’s. Show that: (a) If the ui ’s are independent and identically gamma distributed with f (ui ) = where ui ≥ 0 and θ > 0, then α  OLS − s2 is unbiased for α.

θ−1 −ui 1 e Γ(θ) ui

(b) If the ui ’s are independent and identically χ2 distributed with ν degrees of freedom, then α  OLS − s2 /2 is unbiased for α. (c) If the ui ’s are independent and identically exponentially distributed with f(ui ) = where ui ≥ 0 and θ > 0, then α  OLS − s is consistent for α.

1 −ui /θ θe

11. The Heteroskedastic Consequences of an Arbitrary Variance for the Initial Disturbance of an AR(1) Model. This is based on Baltagi and Li (1990, 1992). Consider a simple AR(1) model ut = ρut−1 + ǫt

t = 1, 2, . . . , T

|ρ| < 1

with ǫt ∼ IID(0, σ 2ǫ ) independent of u0 ∼ (0, σ 2ǫ /τ ), and τ is an arbitrary positive parameter. (a) Show that this arbitrary variance on the initial disturbance u0 renders the disturbances, in general, heteroskedastic. (b) Show that var(ut ) = σ 2t is increasing if τ > (1 − ρ2 ) and decreasing if τ < (1 − ρ2 ). When is the process homoskedastic? (c) Show that cov(ut , ut−s ) = ρs σ 2t−s for t ≥ s. Hint: See the solution by Kim (1991).

(d) Consider the simple regression model yt = βxt + ut

t = 1, 2 . . . , T

with ut following the AR(1) process described above. Consider the common case where ρ > 0 and the xt ’s are positively autocorrelated. For this case, it is a standard result that 2  the var(β OLS ) is understated under the stationary case (i.e., (1 − ρ ) = τ ), see problem 8. This means that OLS rejects too often the hypothesis H0 ; β = 0. Show that OLS will reject more often than the stationary case if τ < 1 − ρ2 and less often than the stationary case if τ > (1 − ρ2 ). Hint: See the solution by Koning (1992). 12. ML Estimation of Linear Regression Model with AR(1) Errors and Two Observations. This is based on Magee (1993). Consider the regression model yi = xi β + ui , with only two observations |x2 | are scalars. Assume that ui ∼ N (0, σ 2 ) and u2 = ρu1 + ǫ i = 1, 2, and the nonstochastic |x1 | = 2 2 with |ρ| < 1. Also, ǫ ∼ N [0, (1 − ρ )σ ] where ǫ and u1 are independent. (a) Show that the OLS estimator of β is (x1 y1 + x2 y2 )/(x21 + x22 ). (b) Show that the ML estimator of β is (x1 y1 − x2 y2 )/(x21 − x22 ).

(c) Show that the ML estimator of ρ is 2x1 x2 /(x21 + x22 ) and thus is nonstochastic.

(d) How do the ML estimates of β and ρ behave as x1 → x2 and x1 → −x2 ? Assume x2 = 0. Hint: See the solution by Baltagi and Li (1995). 13. For the empirical example in section 5.5 based on the Cigarette Consumption Data in Table 3.2.


Problems

123

(a) Replicate the OLS regression of logC on logP , logY and a constant. Plot the residuals versus logY and verify Figure 3.2. (b) Run Glejser’s (1969) test by regressing |ei | the absolute value of the residuals from part (a), on (logYi )δ for δ = 1, −1, −0.5 and 0.5. Verify the t-statistics reported in the text.

(c) Run Goldfeld and Quandt’s (1965) test by ordering the observations according to logYi and omitting 12 central observations. Report the two regressions based on the first and last 17 observations and verify the F -test reported in the text.

(d) Verify the Spearman rank correlation test based on the rank (logYi ) and rank |ei |.

(e) Verify Harvey’s (1976) multiplicative heteroskedasticity test based on regressing loge2i on log(logYi ). (f) Run the Breusch and Pagan (1979) test based on the regression of e2i / σ 2 on logYi , where 46 2 2 σ  = i=1 ei /46.

(g) Run White’s (1980) test for heteroskedasticity.

(h) Run the Jarque and Bera (1987) test for normality of the disturbances.

(i) Compute White’s (1980) heteroskedasticity robust standard errors for the regression in part (a). 14. A Simple Linear Trend Model with AR(1) Disturbances. This is based on Kr¨ amer (1982). (a) Consider the following simple linear trend model Yt = α + β t + ut where ut = ρut−1 + ǫt with |ρ| < 1, ǫt ∼ IID(0, σ 2ǫ ) and var(ut ) = σ 2u = σ 2ǫ /(1 − ρ2 ). Our interest is focused on the estimates of the trend coefficient, β, and the estimators to be considered are OLS, CO (assuming that the true value of ρ is known), the first-difference estimator (FD), and the Generalized Least Squares (GLS), which is Best Linear Unbiased (BLUE) in this case. In the context of the simple linear trend model, the formulas for the variances of these estimators reduce to V (OLS) = 12σ2 {−6ρT +1 [(T − 1)ρ − (T + 1)]2 − (T 3 − T )ρ4 +2(T 2 − 1)(T − 3)ρ3 + 12(T 2 + 1)ρ2 − 2(T 2 − 1)(T + 3)ρ +(T 3 − T )}/(1 − ρ2 )(1 − ρ)4 (T 3 − T )2 V (CO) = 12σ2 (1 − ρ)2 (T 3 − 3T 2 + 2T ), V (F D) = 2σ 2 (1 − ρT −1 )/(1 − ρ2 )(T − 1)2 , V (GLS) = 12σ2 /(T − 1)[(T − 3)(T − 2)ρ2 − 2(T − 3)(T − 1)ρ + T (T + 1)]. (b) Compute these variances and their relative efficiency with respect to the GLS estimator for T = 10, 20, 30, 40 and ρ between −0.9 and 0.9 in 0.1 increments. (c) For a given T , show that the limit of var(OLS)/var(CO) is zero as ρ → 1. Prove that var(F D) and var(GLS) both tend in the limit to σ 2ǫ /(T − 1) < ∞ as ρ → 1. Conclude that var(GLS)/var(F D) tend to 1 as ρ → 1. Also, show that lim [var(GLS)/var(OLS)] = ρ→1

5(T 2 + T )/6(T 2 + 1) < 1 provided T > 3.

(d) For a given ρ, show that var(F D) = O(T −2 ) whereas the variance of the remaining estimators is O(T −3 ). Conclude that lim [var(F D)/var(CO)] = ∞ for any given ρ. T →∞

15. Consider the empirical example in section 5.6, based on the Consumption-Income data in Table 5.3. Obtain this data set from the CONSUMP.DAT file on the Springer web site.


124

Chapter 5: Violations of the Classical Assumptions (a) Replicate the OLS regression of Ct on Yt and a constant, and compute the Durbin-Watson statistic. Test H0 ; ρ = 0 versus H1 ; ρ > 0 at the 5% significance level. (b) Perform the Cochrane-Orcutt procedure and verify the regression results in the text. (c) Perform the two-step Prais-Winsten procedure and verify the regression results in the text. Iterate on the Prais-Winsten procedure. (d) Perform the maximum likelihood procedure and verify the results in the text. (e) Perform Durbin’s regression and verify the results in the text. (f) Test for first-order serial correlation using the Breusch and Godfrey test. (g) Compute the Newey-West heteroskedasticity and autocorrelation-consistent standard errors for the least squares estimates in part (a).

16. Benderly and Zwick (1985) considered the following equation RSt = α + βQt+1 + γPt + ut where RSt = the real return on stocks in year t, Qt+1 = the annual rate of growth of real GNP in year t + 1, and Pt = the rate of inflation in year t. The data is provided on the Springer web site and labeled BENDERLY.ASC. This data covers 31 annual observations for the U.S. over the period 1952-1982. This was obtained from Lott and Ray (1991). This equation is used to test the significance of the inflation rate in explaining real stock returns. Use the sample period 1954-1976 to answer the following questions: (a) Run OLS to estimate the above equation. Remember to use Qt+1 . Is Pt significant in this equation? Plot the residuals against time. Compute the Newey-West heteroskedasticity and autocorrelation-consistent standard errors for these least squares estimates. (b) Test for serial correlation using the D.W. test. (c) Would your decision in (b) change if you used the Breusch-Godfrey test for first-order serial correlation? (d) Run the Cochrane-Orcutt procedure to correct for first-order serial correlation. Report your estimate of ρ. (e) Run a Prais-Winsten procedure accounting for the first observation and report your estimate of ρ. Plot the residuals against time. 17. Using our cross-section Energy/GDP data set in Chapter 3, problem 3.16 consider the following two models: M odel 1: logEn = α + βlogRGDP + u M odel 2: En = α + βRGDP + v Make sure you have corrected the W. Germany observation on EN as described in problem 3.16 part (d). (a) Run OLS on both Models 1 and 2. Test for heteroskedasticity using the Goldfeldt/Quandt Test. Omit c = 6 central observations. Why is heteroskedasticity a problem in Model 2, but not Model 1? (b) For Model 2, test for heteroskedasticity using the Glejser Test. (c) Now use the Breusch-Pagan Test to test for heteroskedasticity on Model 2. (d) Apply White’s Test to Model 2. (e) Do all these tests give the same decision?


Problems

125

(f) Propose and estimate a simple transformation of Model 2, assuming heteroskedasticity of the form σ 2i = σ 2 RGDP 2 . (g) Propose and estimate a simple transformation of Model 2, assuming heteroskedasticity of the form σ 2i = σ 2 (a + bRGDP )2 . (h) Now suppose that heteroskedasticity is of the form σ 2i = σ 2 RGDP γ where γ is an unknown parameter. Propose and estimate a simple transformation for Model 2. Hint: You can write σ 2i as exp{α + γlogRGDP } where α = logσ 2 . (i) Compare the standard errors of the estimates for Model 2 from OLS, also obtain White’s heteroskedasticity-consistent standard errors. Compare them with the simple Weighted Least Squares estimates of the standard errors in parts (f), (g) and (h). What do you conclude?

18. You are given quarterly data from the first quarter of 1965 (1965.1) to the fourth quarter of 1983 (1983.4) on employment in Orange County California (EMP) and real gross national product (RGNP). The data set is in a file called ORANGE.DAT on the Springer web site. (a) Generate the lagged variable of real GNP, call it RGN Pt−1 and estimate the following model by OLS: EM Pt = α + βRGN Pt−1 + ut . (b) What does inspection of the residuals and the Durbin-Watson statistic suggest? (c) Assuming ut = ρut−1 + ǫt where |ρ| < 1 and ǫt ∼ IIN(0, σ 2ǫ ), use the Cochrane-Orcutt procedure to estimate ρ, α and β. Compare the latter estimates and their standard errors with those of OLS. (d) The Cochrane-Orcutt procedure omits the first observation. Perform the Prais-Winsten adjustment. Compare the resulting estimates and standard error with those in part (c). (e) Apply the Breusch-Godfrey test for first and second order autoregression. What do you conclude? (f) Compute the Newey-West heteroskedasticity and autocorrelation-consistent covariance standard errors for the least squares estimates in part (a). 19. Consider the earning data underlying the regression in Table 4.1 and available on the Springer web site as EARN.ASC. (a) Apply White’s test for heteroskedasticity to the regression residuals. (b) Compute White’s heteroskedasticity-consistent standard errors. (c) Test the least squares residuals for normality using the Jarque-Bera test. 20. Harrison and Rubinfield (1978) collected data on 506 census tracts in the Boston area in 1970 to study hedonic housing prices and the willingness to pay for clean air. This data is available on the Springer web site as HEDONIC.XLS. The dependent variable is the Median Value (MV) of owneroccupied homes. The regressors include two structural variables, RM the average number of rooms, and AGE representing the proportion of owner units built prior to 1940. In addition there are eight neighborhood variables: B, the proportion of blacks in the population; LSTAT, the proportion of population that is lower status; CRIM, the crime rate; ZN, the proportion of 25000 square feet residential lots; INDUS, the proportion of nonretail business acres; TAX, the full value property tax rate ($/$10000); PTRATIO, the pupil-teacher ratio; and CHAS represents the dummy variable for Charles River: = 1 if a tract bounds the Charles. There are also two accessibility variables, DIS the weighted distances to five employment centers in the Boston region, and RAD the index of accessibility to radial highways. One more regressor is an air pollution variable NOX, the annual average nitrogen oxide concentration in parts per hundred million. (a) Run OLS of MV on the 13 independent variables and a constant. Plot the residuals.


126

Chapter 5: Violations of the Classical Assumptions (b) Apply White’s tests for heteroskedasticity. (c) Obtain the White heteroskedasticity-consistent standard errors. (d) Test the least squares residuals for normality using the Jarque-Bera test.

References For additional readings consult the econometrics books cited in the Preface. Recent chapters on heteroskedasticity and autocorrelation include Griffiths (2001) and King (2001): Ali, M.M. and C. Giaccotto (1984), “A study of Several New and Existing Tests for Heteroskedasticity in the General Linear Model,” Journal of Econometrics, 26: 355-373. Amemiya, T. (1973), “Regression Analysis When the Variance of the Dependent Variable is Proportional to the Square of its Expectation,” Journal of the American Statistical Association, 68: 928-934. Amemiya, T. (1977), “A Note on a Heteroskedastic Model,” Journal of Econometrics, 6: 365-370. Andrews, D.W.K. (1991), “Heteroskedasticity and Autocorrelation Consistent Covariance Matrix Estimation,” Econometrica, 59: 817-858. Baltagi, B. and Q. Li (1990), “The Heteroskedastic Consequences of an Arbitrary Variance for the Initial Disturbance of an AR(1) Model,” Econometric Theory, Problem 90.3.1, 6: 405. Baltagi, B. and Q. Li (1992), “The Bias of the Standard Errors of OLS for an AR(1) process with an Arbitrary Variance on the Initial Observations,” Econometric Theory, Problem 92.1.4, 8: 146. Baltagi, B. and Q. Li (1995), “ML Estimation of Linear Regression Model with AR(1) Errors and Two Observations,” Econometric Theory, Solution 93.3.2, 11: 641-642. Bartlett’s test, M.S. (1937), “Properties of Sufficiency and Statistical Tests,” Proceedings of the Royal Statistical Society, A, 160: 268-282. Beach, C.M. and J.G. MacKinnon (1978), “A Maximum Likelihood Procedure for Regression with Autocorrelated Errors,” Econometrica, 46: 51-58. Benderly, J. and B. Zwick (1985), “Inflation, Real Balances, Output and Real Stock Returns,” American Economic Review, 75: 1115-1123. Breusch, T.S. (1978), “Testing for Autocorrelation in Dynamic Linear Models,” Australian Economic Papers, 17: 334-355. Breusch, T.S. and A.R. Pagan (1979), “A Simple Test for Heteroskedasticity and Random Coefficient Variation,” Econometrica, 47: 1287-1294. Buse, A. (1984), “Tests for Additive Heteroskedasticity: Goldfeld and Quandt Revisited,” Empirical Economics, 9: 199-216. Carroll, R.H. (1982), “Adapting for Heteroskedasticity in Linear Models,” Annals of Statistics, 10: 12241233. Cochrane, D. and G. Orcutt (1949), “Application of Least Squares Regression to Relationships Containing Autocorrelated Error Terms,” Journal of the American Statistical Association, 44: 32-61. Cragg, J.G. (1992), “Quasi-Aitken Estimation for Heteroskedasticity of Unknown Form,” Journal of Econometrics, 54: 197-202.


References

127

Durbin, J. (1960), “Estimation of Parameters in Time-Series Regression Model,” Journal of the Royal Statistical Society, Series B, 22: 139-153. Durbin, J. and G. Watson (1950), “Testing for Serial Correlation in Least Squares Regression-I,” Biometrika, 37: 409-428. Durbin, J. and G. Watson (1951), “Testing for Serial Correlation in Least Squares Regression-II,” Biometrika, 38: 159-178. Evans, M.A., and M.L. King (1980) “A Further Class of Tests for Heteroskedasticity,” Journal of Econometrics, 37: 265-276. Farebrother, R.W. (1980), “The Durbin-Watson Test for Serial Correlation When There is No Intercept in the Regression,” Econometrica, 48: 1553-1563. Glejser, H. (1969), “A New Test for Heteroskedasticity,” Journal of the American Statistical Association, 64: 316-323. Godfrey, L.G. (1978), “Testing Against General Autoregressive and Moving Average Error Models When the Regressors Include Lagged Dependent Variables,” Econometrica, 46: 1293-1302. Goldfeld, S.M. and R.E. Quandt (1965), “Some Tests for Homoscedasticity,” Journal of the American Statistical Association, 60: 539-547. Goldfeld, S.M. and R.E. Quandt (1972), Nonlinear Methods in Econometrics (North-Holland: Amsterdam). Griffiths, W.E. (2001), “Heteroskedasticity,” Chapter 4 in B.H. Baltagi, (ed.), A Companion to Theoretical Econometrics (Blackwell: Massachusetts). Harrison, M. and B.P. McCabe (1979), “A Test for Heteroskedasticity Based On Ordinary Least Squares Residuals,” Journal of the American Statistical Association, 74: 494-499. Harrison, D. and D.L. Rubinfeld (1978), “Hedonic Housing Prices and the Demand for Clean Air,” Journal of Environmental Economics and Management, 5: 81-102. Harvey, A.C. (1976), “Estimating Regression Models With Multiplicative Heteroskedasticity,” Econometrica, 44: 461-466. Hilderth, C. and J. Lu (1960), “Demand Relations with Autocorrelated Disturbances,” Technical Bulletin 276 (Michigan State University, Agriculture Experiment Station). Jarque, C.M. and A.K. Bera (1987), “A Test for Normality of Observations and Regression Residuals,” International Statistical Review, 55: 163-177. Kim, J.H. (1991), “The Heteroskedastic Consequences of an Arbitrary Variance for the Initial Disturbance of an AR(1) Model,” Econometric Theory, Solution 90.3.1, 7: 544-545. King, M. (2001), “Serial Correlation,” Chapter 2 in B.H. Baltagi, (ed.), A Companion to Theoretical Econometrics (Blackwell: Massachusetts). Koenker, R. (1981), “A Note on Studentizing a Test for Heteroskedasticity,” Journal of Econometrics, 17: 107-112. Koenker, R. and G.W. Bassett, Jr. (1982), “Robust Tests for Heteroskedasticity Based on Regression Quantiles,” Econometrica, 50:43-61. Koning, R.H. (1992), “The Bias of the Standard Errors of OLS for an AR(1) process with an Arbitrary Variance on the Initial Observations,” Econometric Theory, Solution 92.1.4, 9: 149-150.


128

Chapter 5: Violations of the Classical Assumptions

Kr¨ amer, W. (1982), “Note on Estimating Linear Trend When Residuals are Autocorrelated,” Econometrica, 50: 1065-1067. Lott, W.F. and S.C. Ray (1992), Applied Econometrics: Problems With Data Sets (The Dryden Press: New York). Maddala, G.S. (1977), Econometrics (McGraw-Hill: New York). Maeshiro, A. (1976), “Autoregressive Transformation, Trended Independent Variables and Autocorrelated Disturbance Terms,” The Review of Economics and Statistics, 58: 497-500. Maeshiro, A. (1979), “On the Retention of the First Observations in Serial Correlation Adjustment of Regression Models,” International Economic Review, 20: 259-265. Magee L. (1993), “ML Estimation of Linear Regression Model with AR(1) Errors and Two Observations,” Econometric Theory, Problem 93.3.2, 9: 521-522. Mizon, G.E. (1995), “A Simple Message for Autocorrelation Correctors: Don’t,” Journal of Econometrics 69: 267-288. Newey, W.K. and K.D. West (1987), “A Simple, Positive Semi-definite, Heteroskedasticity and Autocorrelation Consistent Covariance Matrix,” Econometrica, 55: 703-708. Oberhofer, W. and J. Kmenta (1974), “A General Procedure for Obtaining Maximum Likelihood Estimates in Generalized Regression Models,” Econometrica, 42: 579-590. Park, R.E. and B.M. Mitchell (1980), “Estimating the Autocorrelated Error Model With Trended Data,” Journal of Econometrics, 13: 185-201. Prais, S. and C. Winsten (1954), “Trend Estimation and Serial Correlation,” Discussion Paper 383 (Cowles Commission: Chicago). Rao, P. and Z. Griliches (1969), “Some Small Sample Properties of Several Two-Stage Regression Methods in the Context of Autocorrelated Errors,” Journal of the American Statistical Association, 64: 253-272. Rao, P. and R. L. Miller (1971), Applied Econometrics (Wadsworth: Belmont). Robinson, P.M. (1987), “Asymptotically Efficient Estimation in the Presence of Heteroskedasticity of Unknown Form,” Econometrica, 55: 875-891. Rutemiller, H.C. and D.A. Bowers (1968), “Estimation in a Heteroskedastic Regression Model,” Journal of the American Statistical Association, 63: 552-557. Savin, N.E. and K.J. White (1977), “The Durbin-Watson Test for Serial Correlation with Extreme Sample Sizes or Many Regressors,” Econometrica, 45: 1989-1996. Szroeter, J. (1978), “A Class of Parametric Tests for Heteroskedasticity in Linear Econometric Models,” Econometrica, 46: 1311-1327. Theil, H. (1978), Introduction to Econometrics (Prentice-Hall: Englewood Cliffs, NJ). Waldman, D.M. (1983), “A Note on Algebraic Equivalence of White’s Test and a Variation of the Godfrey/Breusch-Pagan Test for Heteroskedasticity,” Economics Letters, 13: 197-200. White, H. (1980), “A Heteroskedasticity Consistent Covariance Matrix Estimator and a Direct Test for Heteroskedasticity,” Econometrica, 48: 817-838. Wooldridge, J.M. (1991), “On the Application of Robust, Regression-Based Diagnostics to Models of Conditional Means and Conditional Variances,” Journal of Econometrics, 47: 5-46.


CHAPTER 6

Distributed Lags and Dynamic Models 6.1

Introduction

Many economic models have lagged values of the regressors in the regression equation. For example, it takes time to build roads and highways. Therefore, the effect of this public investment on growth in GNP will show up with a lag, and this effect will probably linger on for several years. It takes time before investment in research and development pays off in new inventions which in turn take time to develop into commercial products. In studying consumption behavior, a change in income may affect consumption over several periods. This is true in the permanent income theory of consumption, where it may take the consumer several periods to determine whether the change in real disposable income was temporary or permanent. For example, is the extra consulting money earned this year going to continue next year? Also, lagged values of real disposable income appear in the regression equation because the consumer takes into account his life time earnings in trying to smooth out his consumption behavior. In turn, one’s life time income may be guessed by looking at past as well as current earnings. In other words, the regression relationship would look like Yt = α + β 0 Xt + β 1 Xt−1 + .. + β s Xt−s + ut

t = 1, 2, . . . , T

(6.1)

where Yt denotes the t-th observation on the dependent variable Y and Xt−s denotes the (t-s)th observation on the independent variable X. α is the intercept and β 0 , β 1 , . . . , β s are the current and lagged coefficients of Xt . Equation (6.1) is known as a distributed lag since it distributes the effect of an increase in income on consumption over s periods. Note that the short-run effect of a unit change in X on Y is given by β o , while the long-run effect of a unit change in X on Y is (β 0 + β 1 + .. + β s ). Suppose that we observe Xt from 1955 to 1995. Xt−1 is the same variable but for the previous period, i.e., 1954-1994. Since 1954 is not observed, we start from 1955 for Xt−1 , and end at 1994. This means that when we lag once, the current Xt series will have to start at 1956 and end at 1995. For practical purposes, this means that when we lag once we loose one observation from the sample. So if we lag s periods, we loose s observations. Furthermore, we are estimating one extra β with every lag. Therefore, there is double jeopardy with respect to loss of degrees of freedom. The number of observations fall (because we are lagging the same series), and the number of parameters to be estimated increase with every lagging variable introduced. Besides the loss of degrees of freedom, the regressors in (6.1) are likely to be highly correlated with each other. In fact most economic time series are usually trended and very highly correlated with their lagged values. This introduces the problem of multicollinearity among the regressors and as we saw in Chapter 4, the higher the multicollinearity among these regressors, the lower is the reliability of the regression estimates. In this model, OLS is still BLUE because the classical assumptions are still satisfied. All we have done in (6.1) is introduce the additional regressors (Xt−1 , . . . , Xt−s ). These regressors are uncorrelated with the disturbances since they are lagged values of Xt , which are by assumption not correlated with ut for every t.


130

Chapter 6: Distributed Lags and Dynamic Models

>E

( s + 1)>

> E 0

1

2

3

s

( s + 1)

Figure 6.1 Linear Distributed Lag

In order to reduce the degrees of freedom problem, one could impose more structure on the β’s. One of the simplest forms imposed on these coefficients is the linear arithmetic lag, (see Figure 6.1), which can be written as β i = [(s + 1) − i]β

for i = 0, 1, . . . , s

(6.2)

The lagged coefficients of X follow a linear distributed lag declining arithmetically from (s+1)β for Xt to β for Xt−s . Substituting (6.2) in (6.1) one gets Yt = α +

s

i=0 β i Xt−i

+ ut = α + β

s

i=0 [(s

+ 1) − i]Xt−i + ut

(6.3)

where the latter equation can be estimated by the regression of Yt on a constant and Zt , where Zt =

s

i=0 [(s

+ 1) − i]Xt−i

This Zt can be calculated given s and Xt . Hence, we have reduced the estimation of β 0 , β 1 , . . . , β s  is obtained, β  can be deduced from (6.2), for i = into the estimation of just one β. Once β i 0, 1, . . . , s. Despite its simplicity, this lag is too restrictive to impose on the regression and is not usually used in practice. Alternatively, one can think of β i = f (i) for i = 0, 1, . . . , s. If f (i) is a continuous function, over a closed interval, then it can be approximated by an r-th degree polynomial, f (i) = a0 + a1 i + . . . + ar ir For example, if r = 2, then β i = a0 + a1 i + a2 i2

for i = 0, 1, 2, . . . , s


6.1

Introduction

131

so that β 0 = a0 β 1 = a0 + a1 + a2 β 2 = a0 + 2a1 + 4a2 .. .. .. . . . β s = a0 + sa1 + s2 a2 Once a0 , a1 , and a2 are estimated, β 0 , β 1 , . . . , β s can be deduced. In fact, substituting β i = a0 + a1 i + a2 i2 in (6.1) we get  (6.4) Yt = α + si=0 (a0 + a1 i + a2 i2 )Xt−i + ut s 2 s s = α + a0 i=0 Xt−i + a1 i=0 iXt−i + a2 i=0 i Xt−i + ut

This last equation, shows that α, a0 ,  a1 and a2 can be estimated from the regression of Yt   on a constant, Z0 = si=0 Xt−i , Z1 = si=0 iXt−i and Z2 = si=0 i2 Xt−i . This procedure was proposed by Almon (1965) and is known as the Almon lag. One of the problems with this procedure is the choice of s and r, the number of lags on Xt , and the degree of the polynomial, respectively. In practice, neither is known. Davidson and MacKinnon (1993) suggest starting with a maximum reasonable lag s∗ that is consistent with the theory and then based on the unrestricted regression, given in (6.1), checking whether the fit of the model deteriorates as s∗ ¯ 2 ; (ii) minimizing is reduced. Some criteria suggested for this choice include: (i) maximizing R Akaike’s (1973) Information Criterion (AIC) with respect to s. This is given by AIC(s) = (RSS/T )e2s/T ; or (iii) minimizing Schwarz (1978) Bayesian Information Criterion (BIC) with respect to s. This is given by BIC(s) = (RSS/T )T s/T where RSS denotes the residual sum ¯ 2 , reward good fit but penalize loss of squares. Note that the AIC and BIC criteria, like R of degrees of freedom associated with a high value of s. These criteria are printed by most regression software including SHAZAM, EViews and SAS. Once the lag length s is chosen it is straight forward to determine r, the degree of the polynomial. Start with a high value of r and construct the Z variables as described s 4 in (6.4). If r = 4 is the highest degree polynomial chosen and a4 , the coefficient of Z4 = i=0 i Xt−4 is insignificant, drop Z4 and run the regression for r = 3. Stop, if the coefficient of Z3 is significant, otherwise drop Z3 and run the regression for r = 2. Applied researchers usually impose end point constraints on this Almon lag. A near end point constraint means that β −1 = 0 in equation (6.1). This means that for equation (6.4), this constraint yields the following restriction on the second degree polynomial in a’s: β −1 = f (−1) = a0 − a1 + a2 = 0. This restriction allows us to solve for a0 given a1 and a2 . In fact, substituting a0 = a1 − a2 into (6.4), the regression becomes Yt = α + a1 (Z1 + Z0 ) + a2 (Z2 − Z0 ) + ut

(6.5)

and once a1 and a2 are estimated, a0 is deduced, and hence the β i ’s. This restriction essentially states that Xt+1 has no effect on Yt . This may not be a plausible assumption, especially in our consumption example, where income next year enters the calculation of permanent income or life time earnings. A more plausible assumption is the far end point constraint, where β s+1 = 0. This means that Xt−(s+1) does not affect Yt . The further you go back in time, the less is the effect on the current period. All we have to be sure of is that we have gone far back enough


132

Chapter 6: Distributed Lags and Dynamic Models

>E

.

. ..

. –1 0 1 2 3

I (I + 1)

E

Figure 6.2 A Polynomial Lag with End Point Constraints

to reach an insignificant effect. This far end point constraint is imposed by removing Xt−(s+1) from the equation as we have done above. But, some researchers impose this restriction on β i = f (i), i.e., by restricting β s+1 = f (s + 1) = 0. This yields for r = 2 the following constraint: a0 +(s+1)a1 +(s+1)2 a2 = 0. Solving for a0 and substituting in (6.4), the constrained regression becomes Yt = α + a1 [Z1 − (s + 1)Z0 ] + a2 [Z2 − (s + 1)2 Z0 ] + ut

(6.6)

One can also impose both end point constraints and reduce the regression into the estimation of one a rather than three a’s. Note that β −1 = β s+1 = 0 can be imposed by not including Xt+1 and Xt−(s+1) in the regression relationship. However, these end point restrictions impose the additional restrictions that the polynomial on which the a’s lie should pass through zero at i = −1 and i = (s + 1), see Figure 6.2. These additional restrictions on the polynomial may not necessarily be true. In other words, the polynomial could intersect the X-axis at points other than −1 or (s + 1). Imposing a restriction, whether true or not, reduces the variance of the estimates, and introduces bias if the restriction is untrue. This is intuitive, because this restriction gives additional information which should increase the reliability of the estimates. The reduction in variance and the introduction of bias naturally lead to Mean Square Error criteria that help determine whether these restrictions should be imposed, see Wallace (1972). These criteria are beyond the scope of this chapter. In general, one should be careful in the use of restrictions that may not be plausible or even valid. In fact, one should always test these restrictions before using them. See Schmidt and Waud (1975). Empirical Example: Using the Consumption-Income data from the Economic Report of the President over the period 1950-1993, given in Table 5.1, we estimate a consumption-income regression imposing a five year lag on income. In this case, all variables are in logs and s = 5 in equation (6.1). Table 6.1 gives the SAS output imposing the linear arithmetic lag given in equation (6.2).  = 0.047 as the coefficient of Yt−5 which is denoted by Note that the SAS output reports β Y LAG5. This is statistically significant with a t-value of 83.1. Note that the coefficient of Yt−4


6.1

Introduction

133

Table 6.1 Regression with Arithmetic Lag Restriction Dependent Variable: C Analysis of Variance Source Model Error C Total

DF 1 37 38

Root MSE Dep Mean C.V.

0.01828 9.14814 0.19978

Sum of Squares 2.30559 0.01236 2.31794

Mean Square 2.30559 0.00033

R-square Adj R-sq

0.9947 0.9945

F Value 6902.791

Prob>F 0.0001

T for H0: Parameter=0 0.873 83.083 83.083 83.083 83.083 83.083 83.083 3.906 0.631 –3.058 –2.857 –1.991

Prob > |T| 0.3880 0.0001 0.0001 0.0001 0.0001 0.0001 0.0001 0.0004 0.5319 0.0041 0.0070 0.0539

Parameter Estimates

Variable INTERCEP Y YLAG1 YLAG2 YLAG3 YLAG4 YLAG5 RESTRICT RESTRICT RESTRICT RESTRICT RESTRICT

DF 1 1 1 1 1 1 1 –1 –1 –1 –1 –1

Parameter Estimate 0.095213 0.280795 0.233996 0.187197 0.140398 0.093598 0.046799 0.007218 0.000781 –0.003911 –0.005374 –0.005208

Standard Error 0.10900175 0.00337969 0.00281641 0.00225313 0.00168985 0.00112656 0.00056328 0.00184780 0.00123799 0.00127903 0.00188105 0.00261513

 and so on. The coefficient of Yt is given by 6β  = 0.281. which is denoted by Y LAG4 is 2β, At the bottom of the regression output, SAS tests each one of these five coefficient restrictions individually. We can see that three of these restrictions are rejected at the 5% level. One can test the arithmetic lag restrictions jointly using an F -test. The Unrestricted Residual Sum of Squares (URSS) is obtained by regressing Ct on Yt , Yt−1 , . . . , Yt−5 and a constant. This yields URSS = 0.00667. The RRSS is given in Table 6.1 as 0.01236 and it involves imposing 5 restrictions given in (6.2). Therefore, F =

(0.01236 − 0.00667)/5 = 5.4597 0.00667/32

and this is distributed as F5,32 under the null hypothesis. The observed F -statistic has a p-value of 0.001 and we reject the linear arithmetic lag restrictions. Next we impose an Almon lag based on a second degree polynomial as described in equation (6.4). Table 6.2 reports the SAS output for s = 5 imposing the near end point constraint. In this  = 0.193, β  = 0.299, . . . , β  = case, the estimated regression coefficients rise and then fall: β 0 1 5  −0.159. Only β 5 is statistically insignificant. In addition, SAS reports a t-test for the near end point restriction which is rejected with a p-value of 0.0001. The Almon lag restrictions can be


134

Chapter 6: Distributed Lags and Dynamic Models

Table 6.2 Almon Polynomial, r = 2, s = 5 and Near End-Point Constraint PDLREG Procedure Dependent Variable = C Ordinary Least Squares Estimates SSE MSE SBC Reg Rsq Durbin-Watson

0.014807 0.000411 –185.504 0.9936 0.6958

DFE Root MSE AIC Total Rsq

36 0.020281 –190.495 0.9936

Variable Intercept Y**0 Y**1 Y**2

DF 1 1 1 1

B Value 0.093289 0.400930 –0.294560 –0.268490

Std Error 0.1241 0.00543 0.0892 0.0492

t Ratio 0.752 73.838 –3.302 –5.459

Approx Prob 0.4571 0.0001 0.0022 0.0001

Restriction Y(–1)

DF –1

L Value 0.005691

Std Error 0.00135

t Ratio 4.203

Approx Prob 0.0001

Variable Y(0) Y(1) Y(2) Y(3) Y(4) Y(5)

Parameter Value 0.19324 0.29859 0.31606 0.24565 0.08735 –0.15883

Std Error 0.027 0.038 0.033 0.012 0.026 0.080

t Ratio 7.16 7.88 9.67 21.20 3.32 –1.99

Approx Prob 0.0001 0.0001 0.0001 0.0001 0.0020 0.0539

Estimate of Lag Distribution −0.159 0 0.3161 | | |  | |  | | |  | | |  | | |  | |             | |

jointly tested using Chow’s F -statistic. The URSS is obtained from the unrestricted regression of Ct on Yt , Yt−1 , . . . , Yt−5 and a constant. This was reported above as URSS = 0.00667. The RRSS, given in Table 6.2, is 0.014807 and involves four restrictions. Therefore,

F =

(0.014807 − 0.00667)/4 = 9.76 0.00667/32

and this is distributed as F4,32 under the null hypothesis. The observed F -statistic has a p-value of 0.00003 and we reject the second degree polynomial Almon lag specification with a near end point constraint. Table 6.3 reports the SAS output for s = 5, imposing the far end point constraint. Note  are decreasing, that this restriction is rejected with a p-value of 0.008. In this case, the β’s  = 0.309, . . . , β  = −0.026 with β  and β  being statistically insignificant. Most  = 0.502, β β 0 1 5 4 5 packages have polynomial distributed lags as part of their standard commands. For example, using EViews, replacing the regressor Y by PDL(Y, 5, 2, 1) indicates the request to fit a five year Almon lag on Y that is of the second-order degree, with a near end point constraint.


6.2

Infinite Distributed Lag

135

Table 6.3 Almon Polynomial, r = 2, s = 5 and Far End-Point Constraint PDLREG Procedure Dependent Variable = C Ordinary Least Squares Estimates SSE MSE SBC Reg Rsq Durbin-Watson

0.009244 0.000257 –203.879 0.9960 0.6372

DFE Root MSE AIC Total Rsq

36 0.016024 –208.87 0.9960

Variable Intercept Y**0 Y**1 Y**2

DF 1 1 1 1

B Value –0.015868 0.405244 –0.441447 –0.133484

Std Error 0.1008 0.00439 0.0706 0.0383

t Ratio –0.157 92.331 –6.255 –3.483

Approx Prob 0.8757 0.0001 0.0001 0.0013

Restriction Y(–1)

DF –1

L Value 0.002758

Std Error 0.00107

t Ratio 2.575

Approx Prob 0.0080

Variable Y(0) Y(1) Y(2) Y(3) Y(4) Y(5)

6.2

Parameter Value 0.50208 0.30916 0.15995 0.05442 –0.00741 –0.02555

Std Error 0.064 0.022 0.008 0.025 0.029 0.021

t Ratio 7.89 14.32 19.82 2.20 –0.26 –1.23

Approx Prob 0.0001 0.0001 0.0001 0.0343 0.7998 0.2268

Estimate of Lag Distribution −0.026 0.5021 | | | | | | | | | | | | | | | | | |

Infinite Distributed Lag

So far we have been dealing with a finite number of lags imposed on Xt . Some lags may be infinite. For example, the investment in building highways and roads several decades ago may still have an effect on today’s growth in GNP. In this case, we write equation (6.1) as  Yt = α + ∞ (6.7) i=0 β i Xt−i + ut t = 1, 2, . . . , T.

There are an infinite number of β i ’s to estimate with only T observations. This can only be feasible if more structure is imposed ∞ on the β i ’s. First, we normalize these β i ’s by their sum, i.e., let wi = β i /β where β = i=0 β i . If all the  β i ’s have the same sign, then the β i ’s take the sign of β and 0 ≤ wi ≤ 1 for all i, with ∞ i=0 wi = 1. This means that the wi ’s can be interpreted as probabilities. In fact, Koyck (1954) imposed the geometric lag on the wi ’s, i.e., wi = (1 − λ)λi for i = 0, 1, . . . , ∞1 . Substituting β i = βwi = β(1 − λ)λi in (6.7) we get Yt = α + β(1 − λ)

∞

i i=0 λ Xt−i

+ ut

(6.8)


136

Chapter 6: Distributed Lags and Dynamic Models

Equation (6.8) is known as the infinite distributed lag form of the Koyck lag. The short-run effect of a unit change  in Xt on Yt isgiven by β(1 − λ); whereas the long-run effect of a unit ∞ change in Xt on Yt is ∞ i=0 β i = β i=0 wi = β. Implicit in the Koyck lag structure is that the effect of a unit change in Xt on Yt declines the further back we go in time. For example, if λ = 1/2, then β 0 = β/2, β 1 = β/4, β 2 = β/8, etc. Defining LXt = Xt−1 , as the lag operator, we have Li Xt = Xt−i , and (6.8) reduces to  i (6.9) Yt = α + β(1 − λ) ∞ i=0 (λL) Xt + ut = α + β(1 − λ)Xt /(1 − λL) + ut ∞ i where we have used the fact that i=0 c = 1/(1 − c). Multiplying the last equation by (1 − λL) one gets Yt − λYt−1 = α(1 − λ) + β(1 − λ)Xt + ut − λut−1 or Yt = λYt−1 + α(1 − λ) + β(1 − λ)Xt + ut − λut−1

(6.10)

This is the autoregressive form of the infinite distributed lag. It is autoregressive because it includes the lagged value of Yt as an explanatory variable. Note that we have reduced the problem of estimating an infinite number of β i ’s into estimating λ and β from (6.10). However, OLS would lead to biased and inconsistent estimates, because (6.10) contains a lagged dependent variable as well as serially correlated errors. In fact the error in (6.10) is a Moving Average process of order one, i.e., MA(1), see Chapter 14. We digress at this stage to give two econometric models which would lead to equations resembling (6.10). 6.2.1

Adaptive Expectations Model (AEM)

Suppose that output Yt is a function of expected sales Xt∗ and that the latter is unobservable, i.e., Yt = α + βXt∗ + ut where expected sales are updated according to the following method ∗ ∗ Xt∗ − Xt−1 = δ(Xt − Xt−1 )

(6.11)

that is, expected sales at time t is a weighted combination of expected sales at time t − 1 and actual sales at time t. In fact, ∗ Xt∗ = δXt + (1 − δ)Xt−1

(6.12)

Equation (6.11) is also an error learning model, where one learns from past experience and adjust expectations after observing current sales. Using the lag operator L, (6.12) can be rewritten as Xt∗ = δXt /[1 − (1 − δ)L]. Substituting this last expression in the above relationship, we get Yt = α + βδXt /[1 − (1 − δ)L] + ut

(6.13)

Multiplying both sides of (6.13) by [1 − (1 − δ)L], we get Yt − (1 − δ)Yt−1 = α[(1 − (1 − δ)] + βδXt + ut − (1 − δ)ut−1 (6.14) looks exactly like (6.10) with λ = (1 − δ).

(6.14)


6.3

6.2.2

Estimation and Testing of Dynamic Models with Serial Correlation

137

Partial Adjustment Model (PAM)

Under this model there is a cost of being out of equilibrium and a cost of adjusting to that equilibrium, i.e., Cost = a(Yt − Yt∗ )2 + b(Yt − Yt−1 )2

(6.15)

where Yt∗ is the target or equilibrium level for Y , whereas Yt is the current level of Y . The first term of (6.15) gives a quadratic loss function proportional to the distance of Yt from the equilibrium level Yt∗ . The second quadratic term represents the cost of adjustment. Minimizing this quadratic cost function with respect to Y , we get Yt = γYt∗ +(1−γ)Yt−1 , where γ = a/(a+b). Note that if the cost of adjustment was zero, then b = 0, γ = 1, and the target is reached immediately. However, there are costs of adjustment, especially in building the desired capital stock. Hence, Yt = γYt∗ + (1 − γ)Yt−1 + ut

(6.16)

where we made this relationship stochastic. If the true relationship is Yt∗ = α + βXt , then from (6.16) Yt = γα + γβXt + (1 − γ)Yt−1 + ut

(6.17)

and this looks like (6.10) with λ = (1 − γ), except for the error term, which is not necessarily MA(1) with the Moving Average parameter λ.

6.3

Estimation and Testing of Dynamic Models with Serial Correlation

Both the AEM and the PAM give equations resembling the autoregressive form of the infinite distributed lag. In all cases, we end up with a lagged dependent variable and an error term that is either Moving Average of order one as in (6.10), or just classical or autoregressive as in (6.17). In this section we study the testing and estimation of such autoregressive or dynamic models. If there is a Yt−1 in the regression equation and the ut ’s are classical disturbances, as may be the case in equation (6.17), then Yt−1 is said to be contemporaneously uncorrelated with the disturbance term ut . In fact, the disturbances satisfy assumptions 1-4 of Chapter 3 and E(Yt−1 ut ) = 0 even though E(Yt−1 ut−1 ) = 0. In other words, Yt−1 is not correlated with the current disturbance ut but it is correlated with the lagged disturbance ut−1 . In this case, as long as the disturbances are not serially correlated, OLS will be biased, but remains consistent and asymptotically efficient. This case is unlikely with economic data given that most macro time-series variables are highly trended. More likely, the ut ’s are serially correlated. In this case, OLS is biased and inconsistent. Intuitively, Yt is related to ut , so Yt−1 is related to ut−1 . If ut and ut−1 are correlated, then Yt−1 and ut are correlated. This means that one of the regressors, lagged Y , is correlated with ut and we have the problem of endogeneity. Let us demonstrate what happens to OLS for the simple autoregressive model with no constant Yt = βYt−1 + ν t

|β| < 1

t = 1, 2, . . . , T

(6.18)


138

Chapter 6: Distributed Lags and Dynamic Models

with ν t = ρν t−1 + ǫt , |ρ| < 1 and ǫt ∼ IIN(0, σ 2ǫ ). One can show, see problem 3, that T T T T 2 2  β OLS = t=2 Yt Yt−1 / t=2 Yt−1 = β + t=2 Yt−1 ν t / t=2 Yt−1

2   with plim(β OLS −β) = asymp. bias(β OLS ) = ρ(1−β )/(1+ρβ). This asymptotic bias is positive if ρ > 0 and negative if ρ < 0. Also, this asymptotic bias can be large for small values of β and large values of ρ. For example, if ρ = 0.9 and β = 0.2, the asymptotic bias for β is 0.73. This is more than 3  times the value  of β.2  Also,  ρ = Tt=2  ν t ν t−1 / Tt=2  ν t = Yt − β ν t−1 where  OLS Yt−1 has

 plim( ρ − ρ) = −ρ(1 − β 2 )/(1 + ρβ) = − asymp.bias(β OLS )

This means that if ρ > 0, then  ρ would be negatively biased. However, if ρ < 0, then  ρ is positively biased. In both cases,  ρ is biased towards zero. In fact, the asymptotic bias of the  D.W. statistic is twice the asymptotic bias of β OLS , see problem 3. This means that the D.W. statistic is biased towards not rejecting the null hypothesis of zero serial correlation. Therefore, if the D.W. statistic rejects the null of ρ = 0, it is doing that when the odds are against it, and therefore confirming our rejection of the null and the presence of serial correlation. If on the other hand it does not reject the null, then the D.W. statistic is uninformative and has to be replaced by another conclusive test for serial correlation. Such an alternative test in the presence of a lagged dependent variable has been developed by Durbin (1970), and the statistic computed is called Durbin’s h. Using (6.10) or (6.17), one computes OLS ignoring its possible bias and  ρ from OLS residuals as shown above. Durbin’s h is given by h= ρ[n/(1 − n var(coeff.  of Yt−1 ))]1/2 .

(6.19)

This is asymptotically distributed N (0, 1) under null hypothesis of ρ = 0. If n[var(coeff.  of Yt−1 )] is greater than one, then h cannot be computed, and Durbin suggests running the OLS residuals et on et−1 and the regressors in the model (including the lagged dependent variable), and testing whether the coefficient of et−1 in this regression is significant. In fact, this test can be generalized to higher order autoregressive errors. Let ut follow an AR(p) process ut = ρ1 ut−1 + ρ2 ut−2 + .. + ρp ut−p + ǫt then this test involves running et on et−1 , et−2 , . . . , et−p and the regressors in the model including Yt−1 . The test statistic for H0 ; ρ1 = ρ2 = .. = ρp = 0; is TR2 which is distributed χ2p . This is the Lagrange multiplier test developed independently by Breusch (1978) and Godfrey (1978) and discussed in Chapter 5. In fact, this test has other useful properties. For example, this test is the same whether the null imposes an AR(p) model or an MA(p) model on the disturbances, see Chapter 14. Kiviet (1986) argues that even though these are large sample tests, the BreuschGodfrey test is preferable to Durbin’s h in small samples. 6.3.1

A Lagged Dependent Variable Model with AR(1) Disturbances

A model with a lagged dependent variable and an autoregressive error term is estimated using instrumental variables (IV). This method will be studied extensively in Chapter 11. In short, the IV method corrects for the correlation between Yt−1 and the error term by replacing Yt−1 with its predicted value Yt−1 . The latter is obtained by regressing Yt−1 on some exogenous variables,


6.3

Estimation and Testing of Dynamic Models with Serial Correlation

139

say a set of Z’s, which are called a set of instruments for Yt−1 . Since these variables are exogenous and uncorrelated with ut , Yt−1 will not be correlated with ut . Suppose the regression equation is Yt = α + βYt−1 + γXt + ut

t = 2, . . . , T

(6.20)

and that at least one exogenous variable Zt exists which will be our instrument for Yt−1 . Regressing Yt−1 on Xt , Zt and a constant, we get Yt−1 = Yt−1 +  νt =  a1 +  a2 Zt +  a3 Xt +  νt.

(6.21)

ν t) Yt = α + β Yt−1 + γXt + (ut + β

(6.22)

Then Yt−1 =  a1 +  a2 Zt +  a3 Xt and is independent of ut , because it is a linear combination of exogenous variables. But, Yt−1 is correlated with ut . This means that  ν t is the part of Yt−1 that is correlated with ut . Substituting Yt−1 = Yt−1 +  ν t in (6.20) we get Yt−1 is uncorrelated with the new error term (ut + β ν t ) because ΣYt−1  ν t = 0 from (6.21). Also, Xt is uncorrelated with ut by assumption. But, from (6.21), Xt also satisfies ΣXt  ν t = 0. Hence, Xt is uncorrelated with the new error term (ut + β ν t ). This means that OLS applied to (6.22) will lead to consistent estimates of α, β and γ. The only remaining question is where do we find instruments like Zt ? This Zt should be (i) uncorrelated with ut , (ii) preferably predicting Yt−1 fairly well, but, not predicting it perfectly, otherwise Yt−1 = Yt−1 . If this happens, we are back to OLS which we know is inconsistent, (iii) Σzt2 /T should be finite and different from zero. Recall ¯ In this case, Xt−1 seems like a natural instrumental variable candidate. It is that zt = Zt − Z. an exogenous variable which would very likely predict Yt−1 fairly well, and satisfies Σx2t−1 /T being finite and different from zero. In other words, (6.21) regresses Yt−1 on a constant, Xt−1 and Xt , and gets Yt−1 . Additional lags on Xt can be used as instruments to improve the small sample properties of this estimator. Substituting Yt−1 in equation (6.22) results in consistent estimates of the regression parameters. Wallis (1967) substituted these consistent estimates in the original equation (6.20) and obtained the residuals u t . Then he computed T T  ρ = [ t=2 u t u t−1 /(T − 1)]/[ t=1 u 2t /T ] + (3/T )

where the last term corrects for the bias in  ρ. At this stage, one can perform a Prais-Winsten procedure on (6.20) using  ρ instead of ρ, see Fomby and Guilkey (1983). An alternative two-step procedure has been proposed by Hatanaka (1974). After estimating (6.22) and obtaining the residuals u t from (6.20), Hatanaka (1974) suggests running Yt∗ = Yt − ∗ ∗ ρYt−2 , Xt = Xt − ρXt−1 and u t−1 . Note that this is the Cochrane-Orcutt  ρYt−1 on Yt−1 = Yt−1 −   2t ignores the t u t−1 / Tt=3 u transformation which ignores the first observation. Also,  ρ = Tt=3 u small sample bias correction factor suggested by Wallis (1967). Let  δ be the coefficient of u t−1 ,   then the efficient estimator of ρ is given by  ρ =  ρ + δ. Hatanaka shows that the resulting estimators are asymptotically equivalent to the MLE in the presence of Normality.

Empirical Example: Consider the Consumption-Income data from the Economic Report of the President over the period 1950-1993 given in Table 5.1. Problem 5 asks the reader to verify that Durbin’s h obtained from the lagged dependent variable model described in (6.20) yields a value of 3.367. This is asymptotically distributed as N (0, 1) under the null hypothesis of no


140

Chapter 6: Distributed Lags and Dynamic Models

serial correlation of the disturbances. This null is soundly rejected. The Bruesch and Godfrey test runs the regression of OLS residuals on their lagged values and the regressors in the model. This yields a T R2 = 7.972. This is distributed as χ21 under the null. Therefore, we reject the hypothesis of no first-order serial correlation. Next, we estimate (6.20) using current and lagged values of income (Yt , Yt−1 and Yt−2 ) as a set of instruments for lagged consumption (Ct−1 ). The regression given by (6.22), yields: t−1 + 0.802 Yt + residuals Ct = −0.053 + 0.196 C (0.1386) (0.0799) (0.1380)

Substituting these estimates in (6.20), one gets the residuals u t . Based on these u t ’s, the Wallis (1967) estimate of ρ yields  ρ = 0.681. Using this  ρ, the Prais-Winsten regression on (6.20) gives the following result: Ct = 0.0007 + 0.169 Ct−1 + 0.822 Yt + residuals (0.007) (0.088) (0.088)

ρ = 0.597 Alternatively, based on u t , one can compute Hatanaka’s (1974) estimate of ρ given by  and run Hatanaka’s regression ∗ + 0.820 Y ∗ + 0.068 u Ct∗ = −0.036 + 0.182 Ct−1 t−1 + residuals t (0.054) (0.098) (0.099) (0.142)

where Ct∗ = Ct −  ρCt−1 . The efficient estimate of ρ is given by   ρ= ρ + 0.068 = 0.665. 6.3.2

A Lagged Dependent Variable Model with MA(1) Disturbances

Zellner and Geisel (1970) estimated the Koyck autoregressive representation of the infinite distributed lag, given in (6.10). In fact, we saw that this could also arise from the AEM, see (6.14). In particular, it is a regression with a lagged dependent variable and an MA(1) error term with the added restriction that the coefficient of Yt−1 is the same as the MA(1) parameter. For simplicity, we write Yt = α + λYt−1 + βXt + (ut − λut−1 )

(6.23)

Let wt = Yt − ut , then (6.23) becomes wt = α + λwt−1 + βXt

(6.24)

By continuous substitution of lagged values of wt in (6.24) we get wt = α(1 + λ + λ2 + .. + λt−1 ) + λt w0 + β(Xt + λXt−1 + .. + λt−1 X1 ) and replacing wt by (Yt − ut ), we get Yt = α(1 + λ + λ2 + .. + λt−1 ) + λt w0 + β(Xt + λXt−1 + .. + λt−1 X1 ) + ut

(6.25)

knowing λ, this equation can be estimated via OLS assuming that the disturbances ut are not serially correlated. Since λ is not known, Zellner and Geisel (1970) suggest a search procedure over λ, where 0 < λ < 1. The regression with the minimum residual sums of squares gives the


6.4

Autoregressive Distributed Lag

141

optimal λ, and the corresponding regression gives the estimates of α, β and w0 . The last coefficient wo = Yo −uo = E(Yo ) can be interpreted as the expected value of the initial observation on the dependent variable. Klein (1958) considered the direct estimation of the infinite Koyck lag, given in (6.8) and arrived at (6.25). The search over λ results in MLEs of the coefficients. Note, however, that the estimate of wo is not consistent. Intuitively, as t tends to infinity, λt tends to zero implying no new information to estimate wo . In fact, some applied researchers ignore the variable λt in the regression given in (6.25). This practice, known as truncating the remainder, is not recommended since the Monte Carlo experiments of Maddala and Rao (1971) and Schmidt (1975) have shown that even for T = 60 or 100, it is not desirable to omit λt from (6.25). In summary, we have learned how to estimate a dynamic model with a lagged dependent variable and serially correlated errors. In case the error is autoregressive of order one, we have outlined the steps to implement the Wallis Two-Stage estimator and Hatanaka’s two-step procedure. In case the error is Moving Average of order one, we have outlined the steps to implement the Zellner-Geisel procedure.

6.4

Autoregressive Distributed Lag

So far, section 6.1 considered finite distributed lags on the explanatory variables, whereas section 6.2 considered an autoregressive relation including the first lag of the dependent variable and current values of the explanatory variables. In general, economic relationships may be generated by an Autoregressive Distributed Lag (ADL) scheme. The simplest form is the ADL (1,1) model which is given by Yt = α + λYt−1 + β 0 Xt + β 1 Xt−1 + ut

(6.26)

where both Yt and Xt are lagged once. By specifying higher order lags for Yt and Xt , say an ADL (p, q) with p lags on Yt and q lags on Xt , one can test whether the specification now is general enough to ensure White noise disturbances. Next, one can test whether some restrictions can be imposed on this general model, like reducing the order of the lags to arrive at a simpler ADL model, or estimating the simpler static model with the Cochrane-Orcutt correction for serial correlation, see problem 20 in Chapter 7. This general to specific modelling strategy is prescribed by David Hendry and is utilized by the econometric software PC-Give, see Gilbert (1986). Returning to the ADL (1, 1) model in (6.26) one can invert the autoregressive form as follows: Yt = α(1 + λ + λ2 + ..) + (1 + λL + λ2 L2 + ..)(β 0 Xt + β 1 Xt−1 + ut )

(6.27)

provided |λ| < 1. This equation gives the effect of a unit change in Xt on future values of Yt . In fact, ∂Yt /∂Xt = β 0 while ∂Yt+1 /∂Xt = β 1 + λβ 0 , etc. This gives the immediate shortrun responses with the long-run effect being the sum of all these partial derivatives yielding (β 0 +β 1 )/(1−λ). This can be alternatively derived from (6.26) at the long-run static equilibrium (Y ∗ , X ∗ ) where Yt = Yt−1 = Y ∗ , Xt = Xt−1 = X ∗ and the disturbance is set equal to zero, i.e., Y∗ =

β + β1 ∗ α + 0 X 1−λ 1−λ

Replacing Yt by Yt−1 + ΔYt and Xt by Xt−1 + ΔXt in (6.26) one gets ΔYt = α + β 0 ΔXt − (1 − λ)Yt−1 + (β 0 + β 1 )Xt−1 + ut

(6.28)


142

Chapter 6: Distributed Lags and Dynamic Models

This can be rewritten as

ΔYt = β 0 ΔXt − (1 − λ) Yt−1 −

β + β1 α − 0 Xt−1 + ut 1−λ 1−λ

(6.29)

Note that the term in brackets contains the long-run equilibrium parameters derived in (6.28). In fact, the term in brackets represents the deviation of Yt−1 from the long-run equilibrium term corresponding to Xt−1 . Equation (6.29) is known as the Error Correction Model (ECM), see Davidson, Hendry, Srba and Yeo (1978). Yt is obtained from Yt−1 by adding the short-run effect of the change in Xt and a long-run equilibrium adjustment term. Since, the disturbances are White noise, this model is estimated by OLS.

Note 1. Other distributions besides the geometric distribution can be considered. In fact, a Pascal distribution was considered by Solow (1960), a rational-lag distribution was considered by Jorgenson (1966), and a Gamma distribution was considered by Schmidt (1974, 1975). See Maddala (1977) for an excellent review.

Problems 1. Consider the Consumption-Income data given in Table 5.1 and provided on the Springer web site as CONSUMP.DAT. Estimate a Consumption-Income regression in logs that allows for a six year lag on income as follows: (a) Use the linear arithmetic lag given in equation (6.2). Show that this result can also be obtained as an Almon lag first-degree polynomial with a far end point constraint. (b) Use an Almon lag second-degree polynomial, described in equation (6.4), imposing the near end point constraint. (c) Use an Almon lag second-degree polynomial imposing the far end point constraint. (d) Use an Almon lag second-degree polynomial imposing both end point constraints. (e) Using Chow’s F -statistic, test the arithmetic lag restrictions given in part (a). (f) Using Chow’s F -statistic, test the Almon lag restrictions implied by the model in part (b). (g) Repeat part (f) for the restrictions imposed in parts (c) and (d). 2. Consider fitting an Almon lag third degree polynomial β i = a0 +a1 i+a2 i2 +a3 i3 for i = 0, 1, . . . , 5, on the Consumption-Income relationship in logarithms. In this case, there are five lags on income, i.e., s = 5. (a) Set up the estimating equation for the ai ’s and report the estimates using OLS.  ) to the (b) What is your estimate of β 3 ? What is the standard error? Can you relate the var(β 3 variances and covariances of the ai ’s?

(c) How would the OLS regression in part (a) change if we impose the near end point constraint β −1 = 0?

(d) Test the near end point constraint.


Problems

143

(e) Test the Almon lag specification given in part (a) against an unrestricted five year lag specification on income. 3. For the simple dynamic model with AR(1) disturbances given in (6.18), 2  (a) Verify that plim(β OLS −β) = ρ(1−β )/(1+ρβ). Hint: From (6.18), Yt−1 = βYt−2 +ν t−1 and ρYt−1 = ρβYt−2 + ρν t−1 . Subtracting this last equation from (6.18) and re-arranging terms, T one gets Yt = (β + ρ)Yt−1 − ρβYt−2 + ǫt . Multiply both sides by Yt−1 and sum t=2 Yt Yt−1 = T    2 2 (β + ρ) t=2 Yt−1 − ρβ Tt=2 Yt−1 Yt−2 + Tt=2 Yt−1 ǫt . Now divide by Tt=2 Yt−1 and take probability limits. See Griliches (1961). (b) For various values of |ρ| < 1 and |β| < 1, tabulate the asymptotic bias computed in part (a).  (c) Verify that plim( ρ − ρ) = −ρ(1 − β 2 )/(1 + ρβ) = −plim(β − β). OLS

(d) Using part (c), show that plim d = 2(1− plim  ρ) = 2[1 −  ν t−1 )2 / Tt=1 ν2t denotes the Durbin-Watson statistic. 

 βρ(β + ρ) ] where d = Tt=2 ( νt − 1 + βρ

T (e) Knowing the true disturbances, the Durbin-Watson statistic would be d∗ = t=2 (ν t −  T ν t−1 )2 / t=1 ν 2t and its plim d∗ = 2(1 − ρ). Using part (d), show that plim (d − d∗ ) = 2ρ(1 − β 2 )  = 2plim(β OLS − β) obtained in part (a). See Nerlove and Wallis (1966). For 1 + βρ various values of |ρ| < 1 and |β| < 1, tabulate d∗ and d and the asymptotic bias in part (d).

4. For the simple dynamic model given in (6.18), let the disturbances follow an MA(1) process ν t = ǫt + θǫt−1 with ǫt ∼ IIN(0, σ 2ǫ ). δ(1 − β 2 ) where δ = θ/(1 + θ2 ). 1 + 2βδ (b) Tabulate this asymptotic bias for various values of |β| < 1 and 0 < θ < 1. 1 T ν2 ) = σ 2ǫ [1 + θ(θ − θ∗ )] where θ∗ = δ(1 − β 2 )/(1 + 2βδ) and  νt = (c) Show that plim( T t=2 t  Yt − β OLS Yt−1 .  (a) Show that plim(β OLS − β) =

5. Consider the lagged dependent variable model given in (6.20). Using the Consumption-Income data from the Economic Report of the President over the period 1950-1993 which is given in Table 5.1. (a) Test for first-order serial correlation in the disturbances using Durbin’s h given in (6.19). (b) Test for first-order serial correlation in the disturbances using the Breusch (1978) and Godfrey (1978) test. (c) Test for second-order serial correlation in the disturbances. 6. Using the U.S. gasoline data in Chapter 4, problem 15 given in Table 4.2 and obtained from the USGAS.ASC file, estimate the following two models:       RGN P CAR QM G = γ 1 + γ 2 log + γ 3 log Static: log CAR t P OP t P OP t  PMG + ǫt +γ 4 log P GN P t Dynamic: log



QM G CAR



   RGN P CAR = γ 1 + γ 2 log + γ3 P OP t t   P OP t   QM G PMG + λlog + ǫt +γ 4 log P GN P t CAR t−1




144

Chapter 6: Distributed Lags and Dynamic Models (a) Compare the implied short-run and long-run elasticities for price (P M G) and income (RGN P ). (b) Compute the elasticities after 3, 5 and 7 years. Do these lags seem plausible? (c) Can you apply the Durbin-Watson test for serial correlation to the dynamic version of this model? Perform Durbin’s h-test for the dynamic gasoline model. Also, the Breusch-Godfrey test for first-order serial correlation.

7. Using the U.S. gasoline data in Chapter 4, problem 15, given in Table 4.2 estimate the following model with a six year lag on prices: log



QM G CAR

   RGN P CAR + γ 3 log P OP t P OP t t    PMG +γ 4 6i=0 wi log P GN P t−i



= γ 1 + γ 2 log



(a) Report the unrestricted OLS estimates. (b) Now, estimate a second degree polynomial lag for the same model. Compare the results with part (a) and explain why you got such different results. (c) Re-estimate part (b) comparing the six year lag to a four year, and eight year lag. Which one would you pick? (d) For the six year lag model, does a third degree polynomial give a better fit? (e) For the model outlined in part (b), reestimate with a far end point constraint. Now, reestimate with only a near end point constraint. Are such restrictions justified in this case?

References This chapter is based on the material in Maddala (1977), Johnston (1984), Kelejian and Oates (1989) and Davidson and MacKinnon (1993). Additional references on the material in this chapter include: Akaike, H. (1973), “Information Theory and an Extension of the Maximum Likelihood Principle,” in B. Petrov and F. Csake, eds. 2nd. International Symposium on Information Theory, Budapest: Akademiai Kiado. Almon, S. (1965), “The Distributed Lag Between Capital Appropriations and Net Expenditures,” Econometrica, 30: 407-423. Breusch, T.S. (1978), “Testing for Autocorrelation in Dynamic Linear Models,” Australian Economic Papers, 17: 334-355. Davidson, J.E.H., D.F. Hendry, F. Srba and S. Yeo (1978), “Econometric Modelling of the Aggregate Time-Series Relationship Between Consumers’ Expenditure and Income in the United Kingdom,” Economic Journal, 88: 661-692. Dhrymes, P.J. (1971), Distributed Lags: Problems of Estimation and Formulation (Holden-Day: San Francisco). Durbin, J. (1970), “Testing for Serial Correlation in Least Squares Regression when Some of the Regressors are Lagged Dependent Variables,” Econometrica, 38: 410-421. Fomby, T.B. and D.K. Guilkey (1983), “An Examination of Two-Step Estimators for Models with Lagged Dependent and Autocorrelated Errors,” Journal of Econometrics, 22: 291-300.


References

145

Gilbert, C.L. (1986), “Professor Hendry’s Econometric Methodology,” Oxford Bulletin of Economics and Statistics, 48: 283-307. Godfrey, L.G. (1978), “Testing Against General Autoregressive and Moving Average Error Models when the Regressors Include Lagged Dependent Variables,” Econometrica, 46: 1293-1302. Griliches, Z. (1961), “A Note on Serial Correlation Bias in Estimates of Distributed Lags,” Econometrica, 29: 65-73. Hatanaka, M. (1974), “An Efficient Two-Step Estimator for the Dynamic Adjustment Model with Autocorrelated Errors,” Journal of Econometrics, 2: 199-220. Jorgenson, D.W. (1966), “Rational Distributed Lag Functions,” Econometrica, 34: 135-149. Kiviet, J.F. (1986), “On The Vigor of Some Misspecification Tests for Modelling Dynamic Relationships,” Review of Economic Studies, 53: 241-262. Klein, L.R. (1958), “The Estimation of Distributed Lags,” Econometrica, 26: 553-565. Koyck, L.M. (1954), Distributed Lags and Investment Analysis (North-Holland: Amsterdam). Maddala, G.S. and A.S. Rao (1971), “Maximum Likelihood Estimation of Solow’s and Jorgenson’s Distributed Lag Models,” Review of Economics and Statistics, 53: 80-88. Nerlove, M. and K.F. Wallis (1967), “Use of the Durbin-Watson Statistic in Inappropriate Situations,” Econometrica, 34: 235-238. Schwarz, G. (1978), “Estimating the Dimension of a Model,” Annals of Statistics, 6: 461-464. Schmidt, P. (1974), “An Argument for the Usefulness of the Gamma Distributed Lag Model,” International Economic Review, 15: 246-250. Schmidt, P. (1975), “The Small Sample Effects of Various Treatments of Truncation Remainders on the Estimation of Distributed Lag Models,” Review of Economics and Statistics, 57: 387-389. Schmidt, P. and R. N. Waud (1973), “The Almon lag Technique and the Monetary versus Fiscal Policy Debate,” Journal of the American Statistical Association, 68: 11-19. Solow, R.M. (1960), “On a Family of Lag Distributions,” Econometrica, 28: 393-406. Wallace, T.D. (1972), “Weaker Criteria and Tests for Linear Restrictions in Regression,” Econometrica, 40: 689-698. Wallis, K.F. (1967), “Lagged Dependent Variables and Serially Correlated Errors: A Reappraisal of Three-Pass Least Squares, ” Review of Economics and Statistics, 49: 555-567. Zellner, A. and M.Geisel (1970), “Analysis of Distributed Lag Models with Application to Consumption Function Estimation,” Econometrica, 38: 865-888.


Part II


CHAPTER 7

The General Linear Model: The Basics 7.1

Introduction

Consider the following regression equation y = Xβ + u where

⎤ ⎡ Y1 X11 X12 . . . X1k ⎢ X21 X22 . . . X2k ⎢ Y2 ⎥ ⎥ ⎢ y=⎢ ⎣ : ⎦;X = ⎣ : : : : Yn Xn1 Xn2 . . . Xnk ⎡

(7.1) ⎤

⎤ ⎡ β1 ⎥ ⎢ ⎥ ⎢ ⎥ ; β = ⎢ β2 ⎥ ; u = ⎢ ⎦ ⎣ : ⎦ ⎣ βk

⎤ u1 u2 ⎥ ⎥ : ⎦ un

with n denoting the number of observations and k the number of variables in the regression, with n > k. In this case, y is a column vector of dimension (n×1) and X is a matrix of dimension (n × k). Each column of X denotes a variable and each row of X denotes an observation on these variables. If y is log(wage) as in the empirical example in Chapter 4, see Table 4.1 then the columns of X contain a column of ones for the constant (usually the first column), weeks worked, years of full time experience, years of education, sex, race, marital status, etc.

7.2

Least Squares Estimation

 Least squares minimizes the residual sum of squares where the residuals are given by e = y −X β  denotes a guess on the regression parameters β. The residual sum of squares and β  RSS = ni=1 e2i = e′ e = (y − Xβ)′ (y − Xβ) = y ′ y − y ′ Xβ − β ′ X ′ y + β ′ X ′ Xβ

The last four terms are scalars as can be verified by their dimensions. It is essential that the reader keep track of the dimensions of the matrices used. This will insure proper multiplication, addition, subtraction of matrices and help the reader obtain the right answers. In fact the middle two terms are the same because the transpose of a scalar is a scalar. For a quick review of some matrix properties, see the Appendix to this chapter. Differentiating the RSS with respect to β one gets ∂RSS/∂β = −2X ′ y + 2X ′ Xβ

(7.2)

where use is made of the following two rules of differentiating matrices. The first is that ∂a′ b/∂b = a and the second is ∂(b′ Ab)/∂b = (A + A′ )b = 2Ab where the last equality holds if A is a symmetric matrix. In the RSS equation a is y ′ X and A is X ′X. The first-order condition for minimization equates the expression in (7.2) to zero. This yields X ′ Xβ = X ′ y

(7.3)


150

Chapter 7: The General Linear Model: The Basics

which is known as the OLS normal equations. As long as X is of full column rank , i.e., of rank k, ′ −1 ′  then X ′ X is nonsingular and the solution to the above equations is β OLS = (X X) X y. Full column rank means that no column of X is a perfect linear combination of the other columns. In other words, no variable in the regression can be obtained from a linear combination of the other variables. Otherwise, at least one of the OLS normal equations becomes redundant. This means that we only have (k − 1) linearly independent equations to solve for k unknown β’s. ′ ′  This yields no solution for β OLS and we say that X X is singular. X X is the sum of squares cross product matrix (SSCP). If it has a column of ones then it will contain the sums, the sum of squares, and the cross-product sum between any two variables ⎡ n  ⎤ . . .  ni=1 Xik n n ni=1 Xi2 n 2 ⎢ ⎥ ... i=1 Xi2 i=1 Xi2 i=1 Xi2 Xik ⎥ X ′X = ⎢ ⎣ ⎦ : : : n n n 2 i=1 Xik i=1 Xik Xi2 . . . i=1 Xik

′ Of course y could be added to this matrix as another variable which will generate X yn and ′ yy automatically pertaining to the variable y will generate i=1 yi , nfor us, i.e., thecolumn n n 2 . To see this, let y X y , and X y , . . . , i=1 i i=1 ik i i=1 i1 i

′ yy y′X ′ Z = [y, X] then Z Z = X ′y X ′X

This matrix summarizes the data and we can compute any regression of one variable in Z on any subset of the remaining variables in Z using only Z ′ Z. Denoting the least squares residuals  by e = y − X β OLS , the OLS normal equations given in (7.3) can be written as ′  X ′ (y − X β OLS ) = X e = 0

(7.4)

Note that if the regression includes a constant, the first column of X will be a vector of ones  and the first equation of (7.4) becomes ni=1 ei = 0. This proves the well known result that if there is a constant in the regression, the OLS residuals sum to zero. Equation (7.4) also indicates that the regressor matrix X is orthogonal to the residuals vector e. This will become clear when we define e in terms of the orthogonal projection matrix on X. This representation allows another interpretation of OLS as a method of moments estimator which was considered in Chapter 2. This follows from the classical assumptions where X satisfies E(X ′ u) = 0. The sample counterpart of this condition yields X ′ e/n = 0. These are the OLS normal equations and therefore, yield the OLS estimates without minimizing the residual sums of squares. See Kelejian and Oates (1989) and the discussion of instrumental variable estimation in Chapter 11. Since data in economics are not generated using experiments like the physical sciences, the X’s are stochastic and we only observe one realization of this data. Consider for example, annual observations for GNP, money supply, unemployment rate, etc. One cannot repeat draws for this data in the real world or fix the X’s to generate new y’s (unless one is performing a Monte Carlo study). So we have to condition on the set of X’s observed, see Chapter 5. Classical Assumptions: u ∼ (0, σ 2 In ) which means that (i) each disturbance ui has zero mean, (ii) constant variance, and (iii) ui and uj for i = j are not correlated. The u’s are known as spherical disturbances. Also, (iv) the conditional expectation of u given X is zero, E(u/X) = 0. Note that the conditioning here is with respect to every regressor in X and for all observations


7.2

Least Squares Estimation

151

i = 1, 2, . . . n. In other words, it is conditional on all the elements of the matrix X. Using (7.1), this implies that E(y/X) = Xβ is linear in β, var(ui /X) = σ 2 and cov(ui , uj /X) = 0. Additionally, we assume that plim X ′ X/n is finite and positive definite and plim X ′ u/n = 0 as n → ∞. Given these classical assumptions, and conditioning on the X’s observed, it is easy to show  that β OLS is unbiased for β. In fact using (7.1) one can write OLS = β + (X ′ X)−1 X ′ u β

(7.5)

Taking expectations, conditioning on the X’s, and using assumptions (i) and (iv), one attains  the unbiasedness result. Furthermore, one can derive the variance-covariance matrix of β OLS from (7.5) since ′ ′ −1 ′ ′ ′ −1    var(β = σ 2 (X ′ X)−1 (7.6) OLS ) = E(β OLS − β)(β OLS − β) = E(X X) X uu X(X X)

this uses assumption (iv) along with the fact that E(uu′ ) = σ 2 In . This variance-covariance  ’s across the diagonal and the pairwise matrix is (k × k) and gives the variances of the β i  off the diagonal. The next theorem shows that among all linear  and β covariances of say β i j  unbiased estimators of c′ β, it is c′ β OLS which has the smallest variance. This is known as the Gauss-Markov Theorem. Theorem 1: Consider the linear estimator a′ y for c′ β, where both a and c are arbitrary vectors  of constants. If a′ y is unbiased for c′ β then var(a′ y) ≥ var(c′ β OLS ).

Proof: For a′ y to be unbiased for c′ β it must follow from (7.1) that E(a′ y) = a′ Xβ + E(a′ u) = a′ Xβ = c′ β which means that a′ X = c′ . Also, var(a′ y) = E(a′ y −c′ β)(a′ y −c′ β)′ = E(a′ uu′ a) = ′ ′ 2 ′  σ 2 a′ a. Comparing this variance with that of c′ β OLS , one gets var(a y)− var(c β OLS ) = σ a a − 2 ′ ′ −1 ′ ′ 2 ′ ′ 2 σ c (X X) c. But, c = a X, therefore this difference becomes σ [a a − a PX a] = σ a′ P¯X a where PX is a projection matrix on the X-plane defined as X(X ′ X)−1 X ′ and P¯X is defined as  In − PX . In fact, PX y = X β  and P¯X y = y − PX y = y − y = e. So that y projects OLS = y the vector y on the X-plane and e is the projection of y on the plane orthogonal to X or perpendicular to X, see Figure 7.1. Both PX and P¯X are idempotent which means that the above difference σ 2 aP¯X a is greater or equal to zero since P¯X is positive semi-definite. To see this, define z = P¯X a, then the above difference is equal to σ 2 z ′ z ≥ 0. The implications of the theorem are important. It means for example, that for the choice of c′ = (1, 0, . . . , 0) one can pick β 1 = c′ β for which the best linear unbiased estimator would 1,OLS = c′ β  OLS . Similarly any β j can be chosen by using c′ = (0, . . . , 1, . . . , 0) which has be β ′  1 in the j-th position and zero elsewhere. Again, the BLUE of β j = c′ β is β j,OLS = c β OLS . k Furthermore, any linear combination of these β’s such as their sum j=1 β j which corresponds   as its BLUE. to c′ = (1, 1, . . . , 1) has the sum k β j=1

j,OLS

The disturbance variance σ 2 is unknown and has to be estimated. Note that E(u′ u) = E(tr(uu′ )) = tr(E(uu′ )) = tr(σ 2 In ) = nσ 2 , so that u′ u/n seems like a natural unbiased estimator for σ 2 . However, u is not observed and is estimated by the OLS residuals e. It is therefore, natural to investigate E(e′ e). In what follows, we show that s2 = e′ e/(n − k) is an unbiased estimator for σ 2 . To prove this, we need the fact that ′ −1 ′  ¯ ¯ e = y − Xβ OLS = y − X(X X) X y = PX y = PX u

(7.7)


152

Chapter 7: The General Linear Model: The Basics

Pxy = e

y

yˆ = P x y

x

Figure 7.1 The Orthogonal Decomposition of y

where the last equality follows from the fact that P¯X X = 0. Hence, E(e′ e) = E(u′ P¯X u) = E(tr{u′ P¯X u}) = E(tr{uu′ P¯X }) = tr(σ 2 P¯X ) = σ 2 tr(P¯X ) = σ 2 (n − k)

where the second equality follows from the fact that the trace of a scalar is a scalar. The third equality from the fact that tr(ABC) = tr(CAB). The fourth equality from the fact that E(trace) = trace{E(.)}, and E(uu′ ) = σ 2 In . The last equality from the fact that tr(P¯X ) = tr(In ) − tr(PX ) = n − tr(X(X ′ X)−1 X ′ )

= n − tr(X ′ X(X ′ X)−1 ) = n − tr(Ik ) = n − k. 2 ′ −1 is given by s2 (X ′ X)−1 .  Hence, an unbiased estimator of var(β OLS ) = σ (X X)  So far we have shown that β OLS is BLUE. It can also be shown that it is consistent for β. In fact, taking probability limits of (7.5) as n → ∞, one gets ′ −1 ′  plim(β OLS ) = plim(β) + plim(X X/n) (X u/n) = β

The first equality uses the fact that the plim of a sum is the sum of the plims. The second equality follows from assumption 1 and the fact that plim of a product is the product of plims.

7.3

Partitioned Regression and the Frisch-Waugh-Lovell Theorem

In Chapter 4, we studied a useful property of least squares which allows us to interpret multiple regression coefficients as simple regression coefficients. This was called the residualing interpretation of multiple regression coefficients. In general, this property applies whenever the k regressors given by X can be separated into two sets of variables X1 and X2 of dimension (n×k1 ) and (n × k2 ) respectively, with X = [X1 , X2 ] and k = k1 + k2 . The regression in equation (7.1) becomes a partitioned regression given by y = Xβ + u = X1 β 1 + X2 β 2 + u

(7.8)


7.3

Partitioned Regression and the Frisch-Waugh-Lovell Theorem

153

One may be interested in the least squares estimates of β 2 corresponding to X2 , but one has to control for the presence of X1 which may include seasonal dummy variables or a time trend, see Frisch and Waugh (1933) and Lovell (1963)1 . The OLS normal equations from (7.8) are as follows: 



′  β X1′ y X1 X1 X1′ X2 1,OLS = (7.9)  X2′ X1 X2′ X2 X2′ y β 2,OLS

These can be solved by partitioned inversion of the matrix on the left, see the Appendix to this chapter, or by solving two equations in two unknowns. Problem 2 asks the reader to verify that −1 ′ ¯ ′ ¯  β 2,OLS = (X2 PX1 X2 ) X2 PX1 y

(7.10)

  ′  −1  ′  β 2,OLS = (X2 X2 ) X2 y

(7.11)

P¯X1 y = P¯X1 X2 β 2 + P¯X1 u

(7.12)

where P¯X1 = In − PX1 and PX1 = X1 (X1′ X1 )−1 X1′ . P¯X1 is the orthogonal projection matrix of X1 and P¯X1 X2 generates the least squares residuals of each column of X2 regressed on all the 2 = P¯X X2 and y = P¯X y, then (7.10) can be written variables in X1 . In fact, if we denote by X 1 1 as  using the fact that P¯X1 is idempotent. This implies that β 2,OLS can be obtained from the  regression of y on X2 . In words, the residuals from regressing y on X1 are in turn regressed upon the residuals from each column of X2 regressed on all the variables in X1 . This was illustrated in Chapter 4 with some examples. Following Davidson and MacKinnon (1993) we denote this result more formally as the Frisch-Waugh-Lovell (FWL) Theorem. In fact, if we premultiply (7.8) by P¯X1 and use the fact that P¯X1 X1 = 0, one gets

The FWL Theorem states that: (1) The least squares estimates of β 2 from equations (7.8) and (7.12) are numerically identical and (2) The least squares residuals from equations (7.8) and (7.12) are identical.  ¯ X is idempotent, it immediately follows that, OLS on (7.12) yields β Using the fact that P 2,OLS 1 as given by equation (7.10). Alternatively, one can start from equation (7.8) and use the result that    ¯ ¯ y = PX y + P¯X y = X β OLS + PX y = X1 β 1,OLS + X2 β 2,OLS + PX y

(7.13)

′ ¯  ¯ X2′ P¯X1 y = X2′ P¯X1 X2 β 2,OLS + X2 PX1 PX y

(7.14)

where PX = X(X ′ X)−1 X ′ and P¯X = In − PX . Premultiplying equation (7.13) by X2′ P¯X1 and using the fact that P¯X1 X1 = 0, one gets

But, PX1 PX = PX1 . Hence, P¯X1 P¯X = P¯X . Using this fact along with P¯X X = P¯X [X1 , X2 ] = 0,  the last term of equation (7.14) drops out yielding the result that β 2,OLS from (7.14) is identical to the expression in (7.10). Note that no partitioned inversion was used in this proof. This proves part (1) of the FWL Theorem.


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Chapter 7: The General Linear Model: The Basics

Also, premultiplying equation (7.13) by P¯X1 and using the fact that P¯X1 P¯X = P¯X , one gets  ¯ P¯X1 y = P¯X1 X2 β 2,OLS + PX y

(7.15)

 Now β 2,OLS was shown to be numerically identical to the least squares estimate obtained from equation (7.12). Hence, the first term on the right hand side of equation (7.15) must be the fitted values from equation (7.12). Since the dependent variables are the same in equations (7.15) and (7.12), P¯X y in equation (7.15) must be the least squares residuals from regression (7.12). But, P¯X y is the least squares residuals from regression (7.8). Hence, the least squares residuals from regressions (7.8) and (7.12) are numerically identical. This proves part (2) of the FWL Theorem. Several applications of the FWL Theorem will be given in this book. Problem 2 shows that if X1 is the vector of ones indicating the presence of a constant in the regression, then regression (7.15) is equivalent to running (yi − y¯) on the set of variables in X2 expressed as deviations from their respective sample means. Problem 3 shows that the FWL Theorem can be used to prove that including a dummy variable for one of the observations in the regression is equivalent to omitting that observation from the regression.

7.4

Maximum Likelihood Estimation

In Chapter 2, we introduced the method of maximum likelihood estimation which is based on specifying the distribution we are sampling from and writing the joint density of our sample. This joint density is then referred to as the likelihood function because it gives for a given set of parameters specifying the distribution, the probability of obtaining the observed sample. See Chapter 2 for several examples. For the regression equation, specifying the distribution of the disturbances in turn specifies the likelihood function. These disturbances could be Poisson, Exponential, Normal, etc. Once this distribution is chosen, the likelihood function is maximized and the MLE of the regression parameters are obtained. Maximum likelihood estimators are desirable because they are (1) consistent under fairly general conditions,2 (2) asymptotically normal, (3) asymptotically efficient and (4) invariant to reparameterizations of the model3 . Some of the undesirable properties of MLE are that (1) it requires explicit distributional assumptions on the disturbances, and (2) their finite sample properties can be quite different from their asymptotic properties. For example, MLE can be biased even though they are consistent, and their covariance estimates can be misleading for small samples. In this section, we derive the MLE under normality of the disturbances. The Normality Assumption: u ∼ N (0, σ 2 In ). This additional assumption allows us to derive distributions of estimators and other random variables. This is important for constructing  confidence intervals and tests of hypotheses. In fact using (7.5) one can easily see that β OLS is a linear combination of the u’s. But, a linear combination of normal random variables is itself  OLS is N (β, σ 2 (X ′ X)−1 ). Similarly y is N (Xβ, σ 2 In ) and a normal random variable. Hence, β 2 ¯ e is N (0, σ PX ). Moreover, we can write the joint probability density function of the u’s as f (u1 , u2 , . . . , un ; β, σ 2 ) = (1/2πσ 2 )n/2 exp(−u′ u/2σ 2 ). To get the likelihood function we make the transformation u = y − Xβ and note that the Jacobian of the transformation is one. Hence f (y1 , y2 , . . . , yn ; β, σ 2 ) = (1/2πσ 2 )n/2 exp{−(y − Xβ)′ (y − Xβ)/2σ 2 }

(7.16)


7.4

Maximum Likelihood Estimation

155

Taking the log of this likelihood, we get logL(β, σ 2 ) = −(n/2)log(2πσ 2 ) − (y − Xβ)′ (y − Xβ)/2σ 2

(7.17)

Maximizing this likelihood with respect to β and σ 2 one gets the maximum likelihood estimators (MLE). Let θ = σ 2 and Q = (y − Xβ)′ (y − Xβ), then 2X ′ y − 2X ′ Xβ ∂logL(β, θ) = ∂β 2θ Q ∂logL(β, θ) n = − 2 ∂θ 2θ 2θ Setting these first-order conditions equal to zero, one gets   β M LE = β OLS

and  θ=σ 2M LE = Q/n = RSS/n = e′ e/n.

Intuitively, only the second term in the log likelihood contains β and that term (without the negative sign) has already been minimized with respect to β in (7.2) giving us the OLS estimator.  M LE is unbiased Note that σ 2M LE differs from s2 only in the degrees of freedom. It is clear that β for β while σ 2M LE is not unbiased for σ 2 . Substituting these MLE’s into (7.17) one gets the maximum value of logL which is  2M LE ) = −(n/2)log(2π σ 2M LE ) − e′ e/2 σ 2M LE logL(β M LE , σ

= −(n/2)log(2π) − (n/2)log(e′ e/n) − n/2

= constant − (n/2)log(e′ e).

In order to get the Cram´er-Rao lower bound for the unbiased estimators of β and σ 2 one first computes the information matrix

2 ∂ logL/∂β∂β ′ ∂ 2 logL/∂β∂σ 2 2 (7.18) I(β, σ ) = −E ∂ 2 logL/∂σ 2 ∂β ′ ∂ 2 logL/∂σ 2 ∂σ 2 Recall, that θ = σ 2 and Q = (y − Xβ)′ (y − Xβ). It is easy to show (see problem 4) that ∂ 2 logL(β, θ) 1 ∂Q = 2 ∂β∂θ 2θ ∂β

and

∂ 2 logL(β, θ) −X ′ (y − Xβ) = ∂θ∂β θ2

Therefore,  2  ∂ logL(β, θ) −E(X ′ u) =0 E = ∂θ∂β θ2 Also ∂ 2 logL(β, θ) −X ′ X = θ ∂β∂β ′

and

−4Q 2n −Q n ∂ 2 logL(β, θ) = 2 3 + 2 = 3 + ∂θ 4θ 4θ θ 2θ 2

so that E



∂ 2 logL(β, θ) ∂θ 2



=

−nθ n −2n + n −n = 2 3 + 2 = 2 θ 2θ 2θ 2θ


156

Chapter 7: The General Linear Model: The Basics

using the fact that E(Q) = nσ 2 = nθ. Hence, I(β, σ 2 ) =

X ′ X/σ 2 0 0 n/2σ 4

(7.19)

The information matrix is block-diagonal between β and σ 2 . This is an important property for regression models with normal disturbances. It implies that the Cram´er-Rao lower bound is I −1 (β, σ 2 ) =

σ 2 (X ′ X)−1 0 0 2σ 4 /n

(7.20)

M LE = β  OLS attains the Cram´er-Rao lower bound. Under normality, β  OLS is Note that β MVU (minimum variance unbiased). This is best among all unbiased estimators not only linear unbiased estimators. By assuming more (in this case normality) we get more (MVU rather than BLUE)4 . Problem 5 derives the variance of s2 under normality of the disturbances. This is found to be 2σ 4 /(n − k). This means that s2 does not attain the Cram´er-Rao lower bound. However, 2  following the theory of complete sufficient statistics one can show that both β OLS and s are MVU for their respective parameters and therefore both are small sample efficient. Note also that σ 2M LE is biased, therefore it is not meaningful to compare its variance to the Cram´er-Rao lower bound. There is a trade-off between bias and variance in estimating σ 2 . Problem 6 looks at all estimators of σ 2 of the type e′ e/r and derives r such that the mean squared error (MSE) is minimized. The choice of r turns out to be (n − k + 2). OLS , now we derive the distribution of s2 . In order to do that We found the distribution of β we need a result from matrix algebra, which is stated without proof, see Graybill (1961). Lemma 1: For every symmetric idempotent matrix A of rank r, there exists an orthogonal matrix P such that P ′ AP = Jr where Jr is a diagonal matrix with the first r elements equal to one and the rest equal to zero. We use this lemma to show that the RSS/σ 2 is a chi-squared with (n − k) degrees of freedom. To see this note that e′ e/σ 2 = u′ P¯X u/σ 2 and that P¯X is symmetric and idempotent of rank (n − k). Using the lemma there exists a matrix P such that P ′ P¯X P = Jn−k is a diagonal matrix with the first (n − k) elements on the diagonal equal to 1 and the last k elements equal to zero. Now make the change of variable v = P ′ u. This makes v ∼ N (0, σ 2 In ) since the v’s are linear combinations of the u’s and P ′ P = In . Replacing u by v in RSS/σ 2 we get  2 2 v ′ P P¯X P v/σ 2 = v ′ Jn−k v/σ 2 = n−k i=1 vi /σ

where the last sum is only over i = 1, 2, . . . , n − k. But, the v’s are independent identically distributed N (0, σ 2 ), hence vi2 /σ 2 is the square of a standardized N (0, 1) random variable which is distributed as a χ21 . Moreover, the sum of independent χ2 random variables is a χ2 random variable with degrees of freedom equal to the sum of the respective degrees of freedom. Hence, RSS/σ 2 is distributed as χ2n−k . The beauty of the above result is that it applies to all quadratic forms u′ Au where A is symmetric and idempotent. We will use this result again in the test of hypotheses section.


7.5

7.5

Prediction

157

Prediction

Let us now predict To periods ahead. Those new observations are assumed to satisfy (7.1). In other words yo = Xo β + uo

(7.21)

What is the Best Linear Unbiased Predictor (BLUP) of E(yo )? From (7.21), E(yo ) = Xo β which OLS is BLUE for is a linear combination of the β’s. Using the Gauss-Markov result yo = Xo β ′ 2  Xo β and the variance of this predictor of E(yo ) is Xo var(β OLS )Xo = σ Xo (X ′ X)−1 Xo′ . But, what if we are interested in the predictor for yo ? The best predictor of uo is zero, so the predictor for yo is still yo but its MSE is ′   E( yo − yo )( yo − yo )′ = E{Xo (β OLS − β) − uo }{Xo (β OLS − β) − uo } ′ 2   = Xo var(β OLS )Xo + σ ITo − 2cov{Xo (β OLS − β), uo } = σ 2 Xo (X ′ X)−1 Xo′ + σ 2 ITo

(7.22)

′ −1 ′  the last equality follows from the fact that (β OLS − β) = (X X) X u and uo have zero covariance. The latter holds because uo and u have zero covariance. Intuitively this says that the future To disturbances are not correlated with the current sample disturbances. Therefore, the predictor of the average consumption of a $20, 000 income household is the same as the predictor of consumption of a specific household whose income is $20, 000. The difference is not in the predictor itself but in the MSE attached to it. The latter MSE being larger. Salkever (1976) suggested a simple way to compute these forecasts and their standard errors. The basic idea is to augment the usual regression in (7.1) with a matrix of observation-specific dummies, i.e., a dummy variable for each period where we want to forecast:

u β X 0 y (7.23) + = uo γ Xo ITo yo or

y ∗ = X ∗ δ + u∗ ′

, γ ′ ).

(7.24) X∗

where δ = (β has in its second part a matrix of dummy variables one for each of the To periods for which we are forecasting. Since these To observations do not serve in the ′ ′ , γ′ ) where estimation, problem 7 asks the reader to verify that OLS on (7.23) yields  δ = (β  In other words, OLS on (7.23) yields the  = (X ′ X)−1 X ′ y,  β γ = yo − yo , and yo = Xo β. OLS estimate of β without the To observations, and the coefficients of the To dummies are the forecast errors. This also means that the first n residuals are the usual OLS residuals  based on the first n observations, whereas the next To residuals are all zero. e = y − Xβ Therefore, s∗2 = s2 = e′ e/(n − k), and the variance covariance matrix of  δ is given by

(X ′ X)−1 (7.25) s2 (X ∗′ X ∗ )−1 = s2 [ITo + Xo (X ′ X)−1 Xo′ ] and the off-diagonal elements are of no interest. This means that the regression package gives  and the estimated variance of the forecast error in one stroke. Note the estimated variance of β that if the forecasts rather than the forecast errors are needed, one can replace yo by zero, and  as required. The variance ITo by −ITo in (7.23). The resulting estimate of γ will be yo = Xo β, of this forecast will be the same as that given in (7.25), see problem 7.


158

Chapter 7: The General Linear Model: The Basics

7.6

Confidence Intervals and Test of Hypotheses

We start by constructing a confidence interval for any linear combination of β, say c′ β. We ′ 2 ′ ′ −1  know that c′ β OLS ∼ N (c β, σ c (X X) c) and it is a scalar. Hence, ′ ′ ′ −1 1/2  zobs = (c′ β OLS − c β)/σ(c (X X) c)

(7.26)

 OLS − c′ β)/s(c′ (X ′ X)−1 c)1/2 tobs = (c′ β

(7.27)

′ ′ −1 1/2  c′ β OLS ± tα/2 s(c (X X) c)

(7.28)

is a standardized N (0, 1) random variable. Replacing σ by s is equivalent to dividing zobs by the square root of a χ2 random variable divided by its degrees of freedom. The latter random variable is (n − k)s2 /σ 2 = RSS/σ 2 which was shown to be a χ2n−k . Problem 8 shows that zobs and RSS/σ 2 are independent. This means that

is a N (0, 1) random variable divided by the square root of an independent χ2n−k /(n − k). This is a t-statistic with (n − k) degrees of freedom. Hence, a 100(1 − α)% confidence interval for c′ β is

Example: Let us say we are predicting one year ahead so that To = 1 and xo is a (1 × k) vector of next year’s observations on the exogenous variables. The 100(1 − α) confidence interval for next year’s forecast of yo will be yo ± tα/2 s(1 + x′o (X ′ X)−1 xo )1/2 . Similarly (7.28) allows us to construct confidence intervals or test any single hypothesis on any single β j (again by picking c to have 1 in its j-th position and zero elsewhere). In this case we get the usual t-statistic reported in any regression package. More importantly, this allows us to test any hypothesis concerning any linear combination of the β’s, e.g., testing that the sum of coefficients of input variables in a Cobb-Douglas production function is equal to one. This is known as a test for constant returns to scale, see Chapter 4.

7.7

Joint Confidence Intervals and Test of Hypotheses

We have learned how to test any single hypothesis involving any linear combination of the β’s. But what if we are interested in testing two or three or more hypotheses involving linear combinations of the β’s. For example, testing that β 2 = β 4 = 0, i.e., that variables X2 and X4 are not significant in the model. This can be written as c′2 β = c′4 β = 0 where c′j is a row vector of zeros with a one in the j-th position. In order to test these two hypotheses simultaneously, we rearrange these restrictions on the β’s in matrix form Rβ = 0 where R′ = [c2 , c4 ]. In a similar fashion, we can rearrange g restrictions on the β’s into this matrix R which will now be of dimension (g × k). Also these restrictions need not be of the form Rβ = 0 and can be of the more general form Rβ = r where r is a (g × 1) vector of constants. For example, β 1 + β 2 = 1 and 3β 3 + 2β 4 = 5 are two such restrictions. Since Rβ is a collection of linear combinations 2 ′ −1 ′  of the β’s, the BLUE of these is Rβ OLS and the latter is distributed N (Rβ, σ R(X X) R ). ′ Standardization of the form encountered with the scalar c β gives us the following: ′ ′ −1 ′ −1 2   (Rβ OLS − Rβ) [R(X X) R ] (Rβ OLS − Rβ)/σ

(7.29)


7.8

Restricted MLE and Restricted Least Squares

159

rather than divide by the variance we multiply by its inverse, and since we divided by the variance rather than the standard deviation we square the numerator which means in vector form premultiplying by its transpose. Problem 9 replaces the matrix R by the vector c′ and shows that (7.29) reduces to the square of the z-statistic observed in (7.26). This also proves that the resulting statistic is distributed as χ21 . But, what is the distribution of (7.29)? The trick is to write it in terms of the original disturbances, i.e., (7.30) u′ X(X ′ X)−1 R′ [R(X ′ X)−1 R]−1 R(X ′ X)−1 X ′ u/σ 2 ′ −1 ′  where (Rβ OLS − Rβ) is replaced by R(X X) X u. Note that (7.30) is quadratic in the disturbances u of the form u′ Au/σ 2 . Problem 10 shows that A is symmetric and idempotent and of rank g. Applying the same proof as given below lemma 1 we get the result that (7.30) is distributed as χ2g . Again σ 2 is unobserved, so we divide by (n − k)s2 /σ 2 which is χ2n−k . This becomes a ratio of two χ2 ’s random variables. If we divide the numerator and denominator χ2 ’s by their respective degrees of freedom and prove that they are independent (see problem 11) the resulting statistic   − r)′ [R(X ′ X)−1 R′ ]−1 (Rβ − r)/gs2 (7.31) (Rβ OLS

OLS

is distributed under the null Rβ = r as an F (g, n − k).

7.8

Restricted MLE and Restricted Least Squares

Maximizing the likelihood function given in (7.16) subject to Rβ = r is equivalent to minimizing the residual sum of squares subject to Rβ = r. Forming the Lagrangian function Ψ(β, μ) = (y − Xβ)′ (y − Xβ) + 2μ′ (Rβ − r)

(7.32)

∂Ψ(β, μ)/∂β = −2X ′ y + 2X ′ Xβ + 2R′ μ = 0

(7.33)

and differentiating with respect to β and μ one gets ∂Ψ(β, μ)/∂μ = 2(Rβ − r) = 0

Solving for μ, we premultiply (7.33) by  μ  = [R(X ′ X)−1 R′ ]−1 (Rβ − r)

R(X ′ X)−1

(7.34) and use (7.34) (7.35)

OLS

Substituting (7.35) in (7.33) we get ′ −1 ′ ′ −1 ′ −1    β RLS = β OLS − (X X) R [R(X X) R ] (Rβ OLS − r)

(7.36)

The restricted least squares estimator of β differs from that of the unrestricted OLS estimator by the second term in (7.36) with the term in parentheses showing the extent to which the  unrestricted OLS estimator satisfies the constraint. Problem 12 shows that β RLS is biased  unless the restriction Rβ = r is satisfied. However, its variance is always less than that of β OLS . This brings in the trade-off between bias and variance and the MSE criteria which was discussed in Chapter 2. The Lagrange Multiplier estimator μ  is distributed N (0, σ 2 [R(X ′ X)−1 R′ ]−1 ) under the null hypothesis. Therefore, to test μ = 0, we use   μ/σ 2 = (Rβ − r)′ [R(X ′ X)−1 R′ ]−1 (Rβ − r)/σ 2 (7.37) μ ′ [R(X ′ X)−1 R′ ] OLS

OLS

Since μ measures the cost of imposing the restriction Rβ = r, it is no surprise that the right hand side of (7.37) was already encountered in (7.29) and is distributed as χ2g .


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Chapter 7: The General Linear Model: The Basics

7.9

Likelihood Ratio, Wald and Lagrange Multiplier Tests

Before we go into the derivations of these three classical tests for the null hypothesis H0 ; Rβ = r, it is important for the reader to review the intuitive graphical explanation of these tests given in Chapter 2. The Likelihood Ratio test of H0 ; Rβ = r is based upon the ratio λ = maxℓr /maxℓu , where maxℓu and maxℓr are the maximum values of the unrestricted and restricted likelihoods, respectively. Let us assume for simplicity that σ 2 is known, then ′ 2   maxℓu = (1/2πσ 2 )n/2 exp{−(y − X β M LE ) (y − X β M LE )/2σ }

  where β M LE = β OLS . Denoting the unrestricted residual sum of squares by URSS, we have maxℓu = (1/2πσ 2 )n/2 exp{−U RSS/2σ 2 }

Similarly, maxℓr is given by ′ 2   maxℓr = (1/2πσ 2 )n/2 exp{−(y − X β RM LE ) (y − X β RM LE )/2σ }

  where β RM LE = β RLS . Denoting the restricted residual sum of squares by RRSS, we have maxℓr = (1/2πσ 2 )n/2 exp{−RRSS /2σ 2 }

Therefore, −2logλ = (RRSS − URSS )/σ 2 . Let us find the relationship between these residual sums of squares.       er = y − X β RLS = y − X β OLS − X(β RLS − β OLS ) = e − X(β RLS − β OLS ) ′ ′     e′r er = e′ e + (β RLS − β OLS ) X X(β RLS − β OLS )

(7.38)

where er denotes the restricted residuals and e′r er the RRSS. The cross-product terms drop out  RLS − β  OLS ) from (7.36) into (7.38), we get: because X ′ e = 0. Substituting the value of (β ′ ′ −1 ′ −1   RRSS − URSS = (Rβ OLS − r) [R(X X) R ] (Rβ OLS − r)

(7.39)

 M LE )I(β  M LE )−1 R(β  M LE )′ ]−1 r(β M LE ) M LE )′ [R(β W = r(β

(7.40)

M LE − r)′ [R(X ′ X)−1 R′ ]−1 (Rβ M LE − r)/σ 2 W = (Rβ

(7.41)

It is now clear that −2logλ is the right hand side of (7.39) divided by σ 2 . In fact, this Likelihood Ratio (LR) statistic is the same as that given in (7.37) and (7.29). Under the null hypothesis Rβ = r , this was shown to be a χ2g . The Wald test of Rβ = r is based upon the unrestricted estimator and the extent of which it satisfies the restriction. More formally, if r(β) = 0 denote the vector of g restrictions on β and ′   R(β M LE ) denotes the (g × k) matrix of partial derivatives ∂r(β)/∂β evaluated at β M LE , then the Wald statistic is given by   where I(β) = −E(∂ 2 logL/∂β∂β ′ ). In this case, r(β) = Rβ − r, R(β M LE ) = R and I(β M LE ) = (X ′ X)/σ 2 as seen in (7.19). Therefore, which is the same as the LR statistic5 .


7.9

Likelihood Ratio, Wald and Lagrange Multiplier Tests

161

The Lagrange Multiplier test is based upon the restricted estimator. In section 7.8, we derived the restricted estimator and the estimated Lagrange Multiplier μ . The Lagrange Multiplier μ is the cost or shadow price of imposing the restrictions Rβ = r. If these restrictions are true, one would expect the estimated Lagrange Multiplier μ  to have mean zero. Therefore, a test for the null hypothesis that μ = 0, is called the LM test and the corresponding test statistic is given in equation (7.37). Alternatively, one can derive the LM test as a score’ test based on the score or the first derivative of the log-likelihood function i.e., S(β) = ∂logL/∂β. The score is zero for the unrestricted MLE, and the score test is based upon the departure of S(β), evaluated at the  restricted estimator β RM LE , from zero. In this case, the score form of the LM statistic is given by ′  −1   LM = S(β RM LE ) I(β RM LE ) S(β RM LE )

(7.42)

For our model, S(β) = (X ′ y − X ′ Xβ)/σ 2 and from equation (7.36) we have

′ 2   S(β RM LE ) = X (y − X β RM LE )/σ OLS + R′ [R(X ′ X)−1 R′ ]−1 (Rβ OLS − r)}/σ 2 = {X ′ y − X ′ X β 2  = R′ [R(X ′ X)−1 R′ ]−1 (Rβ OLS − r)/σ

2 ′ −1  Using (7.20), one gets I −1 (β RM LE ) = σ (X X) . Therefore, the score form of the LM test becomes

LM

′ ′ −1 ′ −1 ′ −1 ′ ′ −1 ′ −1 2   = (Rβ OLS − r) [R(X X) R ] R(X X) R [R(X X) R ] (Rβ OLS − r)/σ ′ ′ −1 ′ −1 2   = (Rβ (7.43) OLS − r) [R(X X) R ] (Rβ OLS − r)/σ

This is numerically identical to the LM test derived in equation (7.37) and to the W and LR ′  /σ 2 from (7.35), so it is clear why the Score  statistics derived above. Note that S(β RM LE ) = R μ and the Lagrangian Multiplier tests are identical. The score form of the LM test can also be obtained as a by-product of an artificial regression.  In fact, S(β) evaluated as β RM LE is given by ′ 2   S(β RM LE ) = X (y − X β RM LE )/σ

 where y − X β RM LE is the vector of restricted residuals. If H0 is true, then this converges asymptotically to u and the asymptotic variance of the vector of scores becomes (X ′ X)/σ 2 . The score test is then based upon ′ ′ −1 ′ 2   (y − X β RM LE ) X(X X) X (y − X β RM LE )/σ

(7.44)

 This expression is the explained sum of squares from the artificial regression of (y −X β RM LE )/σ on X. To see that this is exactly identical to the LM test in equation (7.37), recall from equation ′  on the left hand  (7.33) that R′ μ  = X ′ (y − X β RM LE ) and substituting this expression for R μ 2 side of equation (7.37) we get equation (7.44). In practice, σ is estimated by s2 the Mean Square Error of the restricted regression. This is an example of the Gauss-Newton Regression which will be discussed in Chapter 8. An alternative approach to testing H0 , is to estimate the restricted and unrestricted models and compute the following F -statistic Fobs =

(RRSS − URSS )/g URSS /(n − k)

(7.45)


162

Chapter 7: The General Linear Model: The Basics

This statistic is known in the econometric literature as the Chow (1960) test and was encountered in Chapter 4. Note that from equation (7.39), if we divide the numerator by σ 2 we get a χ2g statistic divided by its degrees of freedom. Also, using the fact that (n − k)s2 /σ 2 is χ2n−k , the denominator divided by σ 2 is a χ2n−k statistic divided by its degrees of freedom. Problem 11 shows independence of the numerator and denominator and completes the proof that Fobs is distributed F (g, n − k) under H0 . Chow’s (1960) Test for Regression Stability Chow (1960) considered the problem of testing the equality of two sets of regression coefficients y1 = X1 β 1 + u1

and

y2 = X2 β 2 + u2

(7.46)

where X1 is n1 × k and X2 is n2 × k with n1 and n2 > k. In this case, the unrestricted regression can be written as

X1 0 β1 u1 y1 = + (7.47) y2 0 X2 β2 u2 under the null hypothesis H0 ; β 1 = β 2 = β, the restricted model becomes

u1 X1 y1 β+ = u2 X2 y2

(7.48)

The URSS and the RRSS are obtained from these two regressions by stacking the n1 + n2 observations. It is easy to show that the URSS = e′1 e1 + e′2 e2 where e1 is the OLS residuals from y1 on X1 and e2 is the OLS residuals from y2 on X2 . In other words, the URSS is the sum of two residual sums of squares from the separate regressions, see problem 13. The Chow F -statistic given in equation (7.45) has k and (n1 + n2 − 2k) degrees of freedom, respectively. Equivalently, one can obtain this Chow F -statistic from running

u1 0 X1 y1 (7.49) (β 2 − β 1 ) + β1 + = u2 X2 X2 y2 Note that the second set of explanatory variables whose coefficients are (β 2 − β 1 ) are interaction variables obtained by multiplying each independent variable in equation (7.48) by a dummy variable, say D2 , that takes on the value 1 if the observation is from the second regression and 0 if it is from the first regression. A test for H0 ; β 1 = β 2 becomes a joint test of significance for the coefficients of these interaction variables. Gujarati (1970) points out that this dummy variable approach has the additional advantage of giving the estimates of (β 2 − β 1 ) and their t-statistics. If the Chow F -test rejects stability, these individual interaction dummy variable coefficients may point to the source of instability. Of course, one has to be careful with the interpretation of these individual t-statistics, after all they can all be insignificant with the joint F -statistic still being significant, see Maddala (1992). In case one of the two regressions does not have sufficient observations to estimate a separate regression say n2 < k, then one can proceed by running the regression on the full data set to get the RRSS. This is the restricted model because the extra n2 observations are assumed to be generated by the same regression as the first n1 observations. The URSS is the residual sums


7.9

Likelihood Ratio, Wald and Lagrange Multiplier Tests

163

of squares based only on the longer period (n1 observations). In this case, the Chow F -statistic given in equation (7.45) has n2 and n1 − k degrees of freedom, respectively. This is known as Chow’s predictive test since it tests whether the shorter n2 observations are different from their predictions using the model with the longer n1 observations. This predictive test can be performed with dummy variables as follows: Introduce n2 observation specific dummies, one for each of the observations in the second regression. Test the joint significance of these n2 dummy variables. Salkever’s (1976) result applies and each dummy variable will have as its estimated coefficient the prediction error with its corresponding standard error and its t-statistic. Once, again, the individual dummies may point out possible outliers, but it is their joint significance that is under question. The W, LR and LM Inequality We have shown that the LR = W = LM for linear restrictions if the log-likelihood is quadratic. However, this is not necessarily the case for more general situations. In fact, in the next chapter where we consider more general variance covariance structure on the disturbances, estimating this variance-covariance matrix destroys this equality and may lead to conflict in hypotheses testing as noted by Berndt and Savin (1977). In this case, W ≥ LR ≥ LM . See also the problems at the end of this chapter. The LR, W and LM tests are based on the efficient MLE. When consistent rather than efficient estimators are used, an alternative way of constructing the score-type test is known as Neyman’s C(α). For details, see Bera and Permaratne (2001). Although, these three tests are asymptotically equivalent, one test may be more convenient than another for a particular problem. For example, when the model is linear but the restriction is nonlinear, the unrestricted model is easier to estimate than the restricted model. So the Wald test suggests itself in that it relies only on the unrestricted estimator. Unfortunately, the Wald test has a drawback that the LR and LM test do not have. In finite samples, the Wald test is not invariant to testing two algebraically equivalent formulations of the nonlinear restriction. This fact has been pointed out in the econometric literature by Gregory and Veall (1985, 1986) and Lafontaine and White (1986). In what follows, we review some of Gregory and Veall’s (1985) findings: Consider the linear regression with two regressors yt = β o + β 1 x1t + β 2 x2t + ut

(7.50)

where the ut ’s are IIN(0, σ 2 ), and the nonlinear restriction β 1 β 2 = 1. Two algebraically equivalent formulation of the null hypothesis are: H A ; r A (β) = β 1 − 1/β 2 = 0, and H B ; r B (β) =  β 1 β 2 − 1 = 0. The unrestricted maximum likelihood estimator is β OLS and the Wald statistic given in (7.40) is ′ ′  −1      W = r(β OLS ) [R(β OLS )V (β OLS )R (β OLS )] r(β OLS )

(7.51)

  where V (β OLS ) is the usual estimated variance-covariance matrix of β OLS . Problem 19 asks the reader to verify that the Wald statistics corresponding to HA and HB using (7.51) are and

2 2  β  2 W A = (β 1 2 − 1) /(β 2 v11 + 2v12 + v22 /β 2 )

2 2  β    2 W B = (β 1 2 − 1) /(β 2 v11 + 2β 1 β 2 v12 + β 1 v22 )

(7.52)

(7.53)


164

Chapter 7: The General Linear Model: The Basics

OLS ) for i, j = 0, 1, 2. These Wald statistics are clearly not where the vij ’s are the elements of V (β identical, and other algebraically equivalent formulations of the null hypothesis can be generated with correspondingly different Wald statistics. Monte Carlo experiments were performed with 1000 replications on the model given in (7.50) with various values for β 1 and β 2 , and for a sample size n = 20, 30, 50, 100, 500. The experiments were run when the null hypothesis is true and when it is false. For n = 20 and β 1 = 10, β 2 = 0.1, so that H0 is satisfied, W A rejects the null when it is true 293 times out of a 1000, while W B rejects the null 65 times out of a 1000. At the 5% level one would expect 50 rejections with a 95% confidence interval [36, 64]. Both W A and W B reject too often but W A performs worse than W B . When n is increased to 500, W A rejects 78 times while W B rejects 39 times out of a 1000. W A still rejects too often although its performance is better than that for n = 20, while W B performs well and is within the 95% confidence region. When n = 20, β 1 = 1 and β 2 = 0.5, so that H0 is not satisfied, W A rejects the null when it is false 65 times out of a 1000 whereas W B rejects it 584 times out of a 1000. For n = 500, both test statistics reject the null in 1000 out of 1000 times. Even in cases where the empirical sizes of the tests appear similar, see Table 1 of Gregory and Veall (1985), in particular the case where β 1 = β 2 = 1, Gregory and Veall find that W A and W B are in conflict about 5% of the time for n = 20, and this conflict drops to 0.5% at n = 500. Problem 20 asks the reader to derive four Wald statistics corresponding to four algebraically equivalent formulations of the common factor restriction analyzed by Hendry and Mizon (1978). Gregory and Veall (1986) give Monte Carlo results on the performance of these Wald statistics for various sample sizes. Once again they find conflict among these tests even when their empirical sizes appear to be similar. Also, the differences among the Wald statistics are much more substantial, and persist even when n is as large as 500. Lafontaine and White (1985) consider a simple regression y = α + βx + γz + u where y is log of per capita consumption of textiles, x is log of per capita real income and z is log of relative prices of textiles, with the data taken from Theil (1971, p.102). The estimated equation is: y = 1.37 + 1.14x − 0.83z (0.31) (0.16) (0.04)

with σ 2 = 0.0001833, and n = 17, with standard errors shown in parentheses. Consider the null hypothesis H0 ; β = 1. Algebraically equivalent formulations of H0 are Hk ; β k = 1 for any exponent k. Applying (7.40) with r(β) = β k − 1 and R(β) = kβ k−1 , one gets the Wald statistic  k − 1)2 /[(kβ  k−1 )2 V (β)]  Wk = (β

(7.54)

 is the OLS estimate of β and V (β)is  its corresponding estimated variance. For every k, where β 2 .05 = 4.6. Wk has a limiting χ1 distribution under H0 . The critical values are χ21,.05 = 3.84 and F1,14 The latter is an exact distribution test for β = 1 under H0 . Lafontaine and White (1985) try  = 1.14 and V (β)  = different integer exponents (±k) where k = 1, 2, 3, 6, 10, 20, 40. Using β (0.16)2 one gets W−20 = 24.56, W1 = 0.84, and W20 = 0.12. The authors conclude that one could get any Wald statistic desired by choosing an appropriate exponent. Since β > 1, Wk is inversely related to k. So, we can find a Wk that exceeds the critical values given by the χ2 and F distributions. In fact, W−20 leads to rejection whereas W1 and W20 do not reject H0 .


Notes

165

For testing nonlinear restrictions, the Wald test is easy to compute. However, it has a serious problem in that it is not invariant to the way the null hypothesis is formulated. In this case, the score test may be difficult to compute, but Neyman’s C(α) test is convenient to use and provide the invariance that is needed, see Dagenais and Dufour (1991).

Notes 1. For example, in a time-series setting, including the time trend in the multiple regression is equivalent to detrending each variable first, by residualing out the effect of time, and then running the regression on these residuals. 2. Two exceptions noted in Davidson and MacKinnon (1993) are the following: One, if the model is not identified asymptotically. For example, yt = β(1/t) + ut for t = 1, 2, . . . , T , will have (1/t) tend to zero as T → ∞. This means that as the sample size increase, there is no information on β. Two, if the number of parameters in the model increase as the sample size increase. For example, the fixed effects model in panel data discussed in Chapter 11.   3. If the MLE of β is β MLE , then the MLE of (1/β) is (1/β MLE ). Note that this invariance property  implies that MLE cannot be in general unbiased. For example, even if β MLE is unbiased for β, by  the above reparameterization, (1/β MLE ) is not unbiased for (1/β).

4. If the distribution of disturbances is not normal, then OLS is still BLUE as long as the assumptions underlying the Gauss-Markov Theorem are satisfied. The MLE in this case will be in general more efficient than OLS as long as the distribution of the errors is correctly specified.  5. Using the Taylor Series approximation of r(β MLE ) around the true parameter vector β, one gets    r(β MLE ) ≃ r(β) + R(β)(β MLE − β). Under the null hypothesis, r(β) = 0 and the var[r(β MLE )] ≃ ′  R(β) var(β MLE )R (β).

Problems 1. Invariance of the Fitted Values and Residuals to Nonsingular Transformations of the Independent Variables. Post-multiply the independent variables in (7.1) by a nonsingular transformation C, so that X ∗ = XC. (a) Show that PX ∗ = PX and P¯X ∗ = P¯X . Conclude that the regression of y on X has the same fitted values and the same residuals as the regression of y on X ∗ . (b) As an application of these results, suppose that every X was multiplied by a constant, say, a change in the units of measurement. Would the fitted values or residuals change when we rerun this regression? (c) Suppose that X contains two regressors X1 and X2 each of dimension n × 1. If we run the regression of y on (X1 − X2 ) and (X1 + X2 ), will this yield the same fitted values and the same residuals as the original regression of y on X1 and X2 ? 2. The FWL Theorem. (a) Using partitioned inverse results from the Appendix, show that the solution to (7.9) yields  β 2,OLS given in (7.10).


166

Chapter 7: The General Linear Model: The Basics   (b) Alternatively, write (7.9) as a system of two equations in two unknowns β 1,OLS and β 2,OLS .  Solve, by eliminating β 1,OLS and show that the resulting solution is given by (7.10).

(c) Using the FWL Theorem, show that if X1 = ιn a vector of ones indicating the presence of  the constant in the regression, and X2 is a set of economic variables, then (i) β 2,OLS can be obtained by running yi − y¯ on the set of variables in X2 expressed as deviations from  their respective sample means. (ii) The least squares estimate of the constant β 1,OLS can ′ ′ ′ ¯ ¯ be retrieved as y¯ − X2 β 2,OLS where X2 = ιn X2 /n is the vector of sample means of the independent variables in X2 .

3. Let y = Xβ + Di γ + u where y is n × 1, X is n × k and Di is a dummy variable that takes the value 1 for the i-th observation and 0 otherwise. Using the FWL Theorem, prove that the least squares ∗′ ∗ −1 ∗′ ∗   X y and γ OLS = yi − x′i β estimates of β and γ from this regression are β OLS = (X X ) OLS , ∗ ∗ where X denotes the X matrix without the i-th observation and y is the y vector without the i-th observation and (yi , x′i ) denotes the i-th observation on the dependent and independent variables. This means that γ OLS is the forecasted OLS residual from the regression of y ∗ on X ∗ for the i-th observation which was essentially excluded from the regression by the inclusion of the dummy variable Di . 4. Maximum Likelihood Estimation. Given the log-likelihood in (7.17),   (a) Derive the first-order conditions for maximization and show that β MLE = β OLS and that σ  2MLE = RSS/n.

(b) Calculate the second derivatives given in (7.18) and verify that the information matrix reduces to (7.19). 5. Given that u ∼ N (0, σ 2 In ), we showed that (n − k)s2 /σ 2 ∼ χ2n−k . Use this fact to prove that, (a) s2 is unbiased for σ 2 . (b) var(s2 ) = 2σ 4 /(n − k). Hint: E(χ2r ) = r and var(χ2r ) = 2r. 6. Consider all estimators of σ 2 of the type σ 2 = e′ e/r = u′ P¯X u/r with u ∼ N (0, σ 2 In ). (a) Find E( σ 2MLE ) and the bias( σ 2MLE ).

(b) Find var( σ 2MLE ) and the MSE( σ 2MLE ).

(c) Compute MSE( σ 2 ) and minimize it with respect to r. Compare with the MSE of s2 and σ  2MLE .

7. Computing Forecasts and Forecast Standard Errors Using a Regression Package. This is based on Salkever (1976). From equations (7.23) and (7.24), show that ′ ′ −1 ′  ′ ′ X y, and γ ′OLS = yo − Xo β ′OLS ) where β (a)  δ OLS = (β OLS = (X X) OLS . Hint: Set up OLS , γ the OLS normal equations and solve two equations in two unknowns. Alternatively, one can use the FWL Theorem to residual out the additional To dummy variables.

(b) e∗OLS = (e′OLS , 0′ )′ and s∗2 = s2 . (c) s∗2 (X ∗′ X ∗ )−1 is given by the expression in (7.25). Hint: Use partitioned inverse.

8.

 (d) Replace yo by 0 and ITo by −ITo in (7.23) and show that γ  = yo = Xo β OLS whereas all the results in parts (a), (b) and (c) remain the same.

 (a) Show that cov(β OLS , e) = 0. (Since both random variables are normally distributed, this proves their independence).


Problems

9.

167

2 ′  (b) Show that β OLS and s are independent. Hint: A linear (Bu) and quadratic (u Au) forms in normal random variables are independent if BA = 0. See Graybill (1961) Theorem 4.17.

(a) Show that if one replaces R by c′ in (7.29) one gets the square of the z-statistic given in (7.26).

(b) Show that when we replace σ 2 by s2 , the χ21 statistic given in part (a) becomes the square of a t-statistic which is distributed as F (1, n − K). Hint: The square of a N (0, 1) is χ21 . Also the ratio of two independent χ2 random variables divided by their degrees of freedom is an F -statistic with these corresponding degrees of freedom, see Chapter 2. 10.

(a) Show that the matrix A defined in (7.30) by u′ Au/σ 2 is symmetric, idempotent and of rank g. (b) Using the same proof given below lemma 1, show that (7.30) is χ2g .

11.

(a) Show that the two quadratic forms s2 = u′ P¯X u/(n − k) and that given in (7.30) are independent. Hint: Two positive semi-definite quadratic forms u′ Au and u′ Bu are independent if and only if AB = 0, see Graybill (1961) Theorem 4.10. (b) Conclude that (7.31) is distributed as an F (g, n − k).

12. Restricted Least Squares.  (a) Show that β RLS given by (7.36) is biased unless Rβ = r. ′ −1 ′  (b) Show that the var(β X u) where RLS ) = var(A(X X) A = IK − (X ′ X)−1 R′ [R(X ′ X)−1 R′ ]−1 R.

Prove that A2 = A, but A′ = A. Conclude that  var(β RLS ) =

σ 2 A(X ′ X)−1 A′ = σ 2 {(X ′ X)−1 −(X ′ X)−1 R′ [R(X ′ X)−1 R′ ]−1 R(X ′ X)−1 }.

  (c) Show that var(β OLS )− var(β RLS ) is a positive semi-definite matrix.

13. The Chow Test.

(a) Show that OLS on (7.47) yields OLS on each equation separately in (7.46). In other words, ′ −1 ′ ′ −1 ′   X1 y1 and β X2 y2 . β 1,OLS = (X1 X1 ) 2,OLS = (X2 X2 )

(b) Show that the residual sum of squares for equation (7.47) is given by RSS1 + RSS2 , where RSSi is the residual sum of squares from running yi on Xi for i = 1, 2.

(c) Show that the Chow F -statistic can be obtained from (7.49) by testing for the joint significance of Ho ; β 2 − β 1 = 0. 14. Suppose we would like to test Ho ; β 2 = 0 in the following unrestricted model given also in (7.8) y = Xβ + u = X1 β 1 + X2 β 2 + u (a) Using the FWL Theorem, show that the URSS is identical to the residual sum of squares obtained from P¯X1 y = P¯X1 X2 β 2 + P¯X1 u. Conclude that URSS = y ′ P¯X y = y ′ P¯X1 y − y ′ P¯X1 X2 (X2′ P¯X1 X2 )−1 X2′ P¯X1 y. (b) Show that the numerator of the F -statistic for testing Ho ; β 2 = 0 which is given in (7.45), is y ′ P¯X1 X2 (X2′ P¯X1 X2 )−1 X2′ P¯X1 y/k2 . Substituting y = X1 β 1 + u under the null hypothesis, show that the above expression reduces to u′ P¯X1 X2 (X2′ P¯X1 X2 )−1 X2′ P¯X1 u/k2 .


168

Chapter 7: The General Linear Model: The Basics (c) Let v = X2′ P¯X1 u, show that if u ∼ IIN(0, σ 2 ) then v ∼ N (0, σ 2 X2′ P¯X1 X2 ). Conclude that the numerator of the F -statistic given in part (b) when divided by σ 2 can be written as v ′ [var(v)]−1 v/k2 where v ′ [var(v)]−1 v is distributed as χ2k2 under Ho . Hint: See the discussion below lemma 1. (d) Using the result that (n − k)s2 /σ 2 ∼ χ2n−k where s2 is the URSS /(n − k), show that the F -statistic given by (7.45) is distributed as F (k2 , n − k) under Ho . Hint: You need to show that u′ P¯X u is independent of the quadratic term given in part (b), see problem 11. (e) Show that the Wald Test for Ho ; β 2 = 0, given in (7.41), reduces in this case to W =  ′ [R(X ′ X)−1 R′ ]−1 β  denotes the OLS or equivalently the MLE  /s2 were R = [0, Ik ], β β 2 2 2 2 2 of β 2 from the unrestricted model and s is the corresponding estimate of σ 2 given by URSS /(n − k). Using partitioned inversion or the FWL Theorem, show that the numerator of W is k2 times the expression in part (b). (f) Show that the score form of the LM statistic, given in (7.42) and (7.44), can be obtained as the explained sum of squares from the artificial regression of the restricted residuals   on the matrix of regressors X. In this case, s2 = RRSS /(n − k1) (y − X1 β 1,RLS ) deflated by s is the Mean Square Error of the restricted regression. In other words, obtain the explained sum of squares from regressing P¯X1 y/ s on X1 and X2 .

15. Iterative Estimation in Partitioned Regression Models. This is based on Fiebig (1995). Consider the partitioned regression model given in (7.8) and let X2 be a single regressor, call it x2 of dimension n × 1 so that β 2 is a scalar. Consider the following strategy for estimating β 2 : Estimate β 1 from (1) the shortened regression of y on X1 . Regress the residuals from this regression on x2 to yield b2 . (1)

(a) Prove that b2 is biased. Now consider the following iterative strategy for re-estimating β 2 : (1)

Re-estimate β 1 by regressing y − x2 b2 following scheme:

(1)

on X1 to yield b1 . Next iterate according to the

(j)

(j)

b1 = (X1′ X1 )−1 X1′ (y − x2 b2 ) (j+1)

b2

(j)

= (x′2 x2 )−1 x′2 (y − X1 b1 ), (j+1)

(b) Determine the behavior of the bias of b2 (c) Show that as j increases on (7.8).

(j+1) b2

j = 1, 2, ... as j increases.

converges to the estimator of β 2 obtained by running OLS

16. Maddala (1992, pp. 120-127). Consider the simple linear regression yi = α + βXi + ui

i = 1, 2, . . . , n.

where α and β are scalars and ui ∼ IIN(0, σ 2 ). For Ho ; β = 0, (a) Derive the Likelihood Ratio (LR) statistic and show that it can be written as nlog[1/(1 − r2)] where r2 is the square of the correlation coefficient between X and y. as nr2 /(1 − (b) Derive the Wald (W) statistic for testing Ho ; β = 0. Show that it can  nbe written 2 2 2 r ). This is the square of the usual t-statistic on β with σ  MLE = i=1 ei /n used instead of  is the unrestricted MLE which is OLS in this case, and the ei ’s are s2 in estimating σ 2 . β the usual least squares residuals. (c) Derive the Lagrange Multiplier (LM) statistic for testing Ho ; β = 0. Show that it can be 2RMLE = ni=1 (yi − y¯)2 /n written as nr2 . This is the square of the usual t-statistic on with σ  2RMLE is restricted MLE of σ 2 (i.e., imposing Ho used instead of s2 in estimating σ 2 . The σ and maximizing the likelihood with respect to σ 2 ).


Problems

169

(d) Show that LM/n = (W/n)/[1 + (W/n)], and LR/n = log[1 + (W/n)]. Using the following inequality x ≥ log(1 + x) ≥ x/(1 + x), conclude that W ≥ LR ≥ LM . Hint: Use x = W/n.

(e) For the cigarette consumption data given in Table 3.2, compute the W, LR, LM for the simple regression of logC on logP and demonstrate the above inequality given in part (d) for testing that the price elasticity is zero?

17. Engle (1984, pp.785-786). Consider a set of T independent observations on a Bernoulli random variable which takes on the values yt = 1 with probability θ, and yt = 0 with probability (1 − θ). (a) Derive the log-likelihood function, the MLE of θ, the score S(θ), and the information I(θ). (b) Compute the LR, W and LM test statistics for testing Ho ; θ = θ o , versus HA ; θ = θo for θǫ(0, 1). 18. Engle (1984, pp. 787-788). Consider the linear regression model y = Xβ + u = X1 β 1 + X2 β 2 + u given in (7.8), where u ∼ N (0, σ2 IT ). (a) Write down the log-likelihood function, find the MLE of β and σ 2 . (b) Write down the score S(β) and show that the information matrix is block-diagonal between β and σ 2 . (c) Derive the W, LR and LM test statistics in order to test Ho ; β 1 = β o1 , versus HA ; β 1 = β o1 , where β 1 is say the first k1 elements of β. Show that if X = [X1 , X2 ], then 2  )/  )′ [X ′ P¯X X1 ](β o − β W = (β o1 − β 1 1 σ 1 1 2 LM = u ′ X1 [X1′ P¯X2 X1 ]−1 X1′ u / σ2

LR = T log( u′ u / u′ u )

 u  and σ  is the unrestricted MLE, ′ u /T , σ 2 = u ′ u /T . β where u  = y − X β,  = y − Xβ 2 = u  whereas β is the restricted MLE.

(d) Using the above results, show that W = T ( u′ u −u ′ u )/ u′ u 

LM = T ( u′ u −u ′ u )/ u′ u 

Also, that LR = T log[1 + (W/T )]; LM = W/[1 + (W/T )]; and (T − k)W/T k1 ∼ Fk1 ,T −k under Ho . As in problem 16, we use the inequality x ≥ log(1 + x) ≥ x/(1 + x) to conclude that W ≥ LR ≥ LM. Hint: Use x = W/T . However, it is important to note that all the test statistics are monotonic functions of the F -statistic and exact tests for each would produce identical critical regions. (e) For the cigarette consumption data given in Table 3.2, run the following regression: logC = α + βlogP + γlogY + u compute the W, LR, LM given in part (c) for the null hypothesis Ho ; β = −1.

(f) Compute the Wald statistics for HoA ; β = −1, HoB ; β 5 = −1 and HoC ; β −5 = −1. How do these statistics compare? 19. Gregory and Veall (1985). Using equation (7.51) and the two formulations of the null hypothesis H A and H B given below (7.50), verify that the Wald statistics corresponding to these two formulations are those given in (7.52) and (7.53), respectively.


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Chapter 7: The General Linear Model: The Basics

20. Gregory and Veall (1986). Consider the dynamic equation yt = ρyt−1 + β 1 xt + β 2 xt−1 + ut where |ρ| < 1, and ut ∼ NID(0, σ 2 ). Note that for this equation to be the Cochrane-Orcutt transformation yt − ρyt−1 = β 1 (xt − ρxt−1 ) + ut the following nonlinear restriction must be satisfied −β 1 ρ = β 2 called the common factor restriction by Hendry and Mizon (1978). Now consider the following four formulations of this restriction H A ; β 1 ρ + β 2 = 0; H B ; β 1 + (β 2 /ρ) = 0; H C ; ρ + (β 2 /β 1 ) = 0 and H D ; (β 1 ρ/β 2 ) + 1 = 0. (a) Using equation (7.51) derive the four Wald statistics corresponding to the four formulations of the null hypothesis. (b) Apply these four Wald statistics to the equation relating real personal consumption expenditures to real disposable personal income in the U.S. over the post World War II period 1950-1993, see Table 5.1. 21. Effect of Additional Regressors on R2 . This problem was considered in non-matrix form in Chapter 4, problem 4. Regress y on X1 which is T × K1 and compute SSE1 . Add X2 which is T × K2 so that the number of regressors in now K = K1 + K2 . Regress y on X = [X1 , X2 ] and get SSE2 . Show that SSE2 ≤ SSE1 . Conclude that the corresponding R-squares satisfy R22 ≥ R12 . Hint: Show that PX − PX1 is a positive semi-definite matrix.

References Additional readings for the material covered in this chapter can be found in Davidson and MacKinnon (1993), Kelejian and Oates (1989), Maddala (1992), Fomby, Hill and Johnson (1984), Greene (1993), Johnston (1984), Judge et al. (1985) and Theil (1971). These econometrics texts were cited earlier. Other references cited in this chapter are the following: Bera A.K. and G. Permaratne (2001), “General Hypothesis Testing,” Chapter 2 in Baltagi, B.H. (ed.), A Companion to Theoretical Econometrics (Blackwell: Massachusetts). Berndt, E.R. and N.E. Savin (1977), “Conflict Among Criteria for Testing Hypotheses in the Multivariate Linear Regression Model,” Econometrica, 45: 1263-1278. Buse, A.(1982), “The Likelihood Ratio, Wald, and Lagrange Multiplier Tests: An Expository Note,” The American Statistician, 36: 153-157. Chow, G.C. (1960), “Tests of Equality Between Sets of Coefficients in Two Linear Regressions,” Econometrica, 28: 591-605. Dagenais, M.G. and J. M. Dufour (1991), “Invariance, Nonlinear Models, and Asymptotic Tests,” Econometrica, 59: 1601-1615. Engle, R.F. (1984), “Wald, Likelihood Ratio, and Lagrange Multiplier Tests in Econometrics,” In: Griliches, Z. and M.D. Intrilligator (eds) Handbook of Econometrics (North-Holland: Amsterdam). Fiebig, D.G. (1995), “Iterative Estimation in Partitioned Regression Models,” Econometric Theory, Problem 95.5.1, 11:1177.


Appendix

171

Frisch, R., and F.V. Waugh (1933), “Partial Time Regression as Compared with Individual Trends,” Econometrica, 1: 387-401. Graybill, F.A.(1961), An Introduction to Linear Statistical Models, Vol. 1 (McGraw-Hill: New York). Gregory, A.W. and M.R. Veall (1985), “Formulating Wald Tests of Nonlinear Restrictions,” Econometrica, 53: 1465-1468. Gregory, A.W. and M.R. Veall (1986), “Wald Tests of Common Factor Restrictions,” Economics Letters, 22: 203-208. Gujarati, D. (1970), “Use of Dummy Variables in Testing for Equality Between Sets of Coefficients in Two Linear Regressions: A Generalization,” The American Statistician, 24: 50-52. Hendry, D.F. and G.E. Mizon (1978), “Serial Correlation as a Convenient Simplification, Not as a Nuisance: A Comment on A Study of the Demand for Money by the Bank of England,” Economic Journal, 88: 549-563. Lafontaine, F. and K.J. White (1986), “Obtaining Any Wald Statistic You Want,” Economics Letters, 21: 35-40. Lovell, M.C. (1963), “Seasonal Adjustment of Economic Time Series,” Journal of the American Statistical Association, 58: 993-1010. Salkever, D. (1976), “The Use of Dummy Variables to Compute Predictions, Prediction Errors, and Confidence Intervals,” Journal of Econometrics, 4: 393-397.

Appendix Some Useful Matrix Properties This book assumes that the reader has encountered matrices before, and knows how to add, subtract and multiply conformable matrices. In addition, that the reader is familiar with the transpose, trace, rank, determinant and inverse of a matrix. Unfamiliar readers should consult standard texts like Bellman (1970) or Searle (1982). The purpose of this Appendix is to review some useful matrix properties that are used in the text and provide easy access to these properties. Most of these properties are given without proof. Starting with Chapter 7, our data matrix X is organized such that it has n rows and k columns, so that each row denotes an observation on k variables and each column denotes n observations on one variable. This matrix is of dimension n × k. The rank of an n × k matrix is always less than or equal to its smaller dimension. Since n > k, the rank (X) ≤ k. When there is no perfect multicollinearity among the variables in X, this matrix is said to be of full column rank k. In this case, X ′ X, the matrix of cross-products is of dimension k × k. It is square, symmetric and of full rank k. This uses the fact that the rank(X ′ X) = rank(X) = k. Therefore, (X ′ X) is nonsingular and the inverse (X ′ X)−1 exists. This is needed for the computation of Ordinary Least Squares. In fact, for least squares to be feasible, X should be of full column rank k and no variable in X should be a perfect linear combination of the other variables in X. If we write ⎡ ′ ⎤ x1 ⎥ ⎢ X = ⎣ ... ⎦ x′n

where x′i denotes the i-th observation, in the data, then X ′ X = of dimension k × 1.

n

i=1

xi x′i where xi is a column vector


172

Chapter 7: The General Linear Model: The Basics

An important and widely encountered matrix is the Identity matrix which will be denoted by In and subscripted by its dimension n. This is a square n×n matrix whose diagonal elements are all equal to one and its off diagonal elements are all equal to zero. Also, σ 2 In will be a familiar scalar covariance matrix, with every diagonal element equal to σ 2 reflecting homoskedasticity or equal variances (see Chapter 5), and zero covariances or no serial correlation (see Chapter 5). Let ⎡ 2 ⎤ 0 σ1 ⎢ ⎥ .. Ω = diag[σ 2i ] = ⎣ ⎦ . 0

σ 2n

2, . . . , n. This matrix be an (n × n) diagonal matrix with the i-th diagonal element equal to σ 2i for i = 1, n will be encountered under heteroskedasticity, see Chapter 9. Note that tr(Ω) = i=1 σ 2i is the sum of its diagonal elements. Also, tr(In ) = n and tr(σ 2 In ) = nσ 2 . Another useful matrix is the projection matrix PX = X(X ′ X)−1 X ′ which is of dimension n × n. This matrix is encountered in Chapter 7. If y denotes the n × 1 vector of observations on the dependent variable, then PX y generates the predicted values y from the least squares regression of y on X. This matrix PX is symmetric and idempotent. This means ′ 2 that PX = PX and PX = PX PX = PX as can be easily verified. Some of the properties of idempotent matrices is that their rank is equal to their trace. Hence, rank(PX ) = tr(PX ) = tr[X(X ′ X)−1 X ′ ] = tr[X ′ X(X ′ X −1 )] = tr(Ik ) = k. Here, we used the fact that tr(ABC) = tr(CAB) = tr(BCA). In other words, the trace is unaffected by the cyclical permutation of the product. Of course, these matrices should be conformable and the product should result in a square matrix. Note that P¯X = In − PX is also a symmetric and idempotent matrix.  In this case, P¯X y = y − PX y = y − y = e where e denotes the least squares residuals, y − X β OLS where ′ −1 ′  β OLS = (X X) X y, see Chapter 7. Some properties of these projection matrices are the following: PX X = X, P¯X X = 0, P¯X e = e

and PX e = 0.

In fact, X ′ e = 0 means that the matrix X is orthogonal to the vector of least squares residuals e. Note ′ ′   that X ′ e = 0 means that X ′ (y − X β OLS ) = 0 or X y = X X β OLS . These k equations are known as the  OLS normal equations and their solution yields the least squares estimates β OLS . By the definition of ¯ ¯ ¯ PX , we have (i) PX + PX = In . Also, (ii) PX and PX are idempotent and (iii) PX P¯X = 0. In fact, any two of these properties imply the third. The rank(P¯X ) = tr(P¯X ) = tr(In − PX ) = n − k. Note that PX and P¯X are of rank k and (n − k), respectively. Both matrices are not of full column rank. In fact, the only full rank, symmetric idempotent matrix is the identity matrix. Matrices not of full rank are singular, and their inverse do not exist. However, one can find a generalized inverse of a matrix Ω which we will call Ω− which satisfies the following requirements: (i) ΩΩ− Ω =Ω (iii) Ω− Ω is symmetric

and

(ii) Ω− ΩΩ− = Ω− (iv) ΩΩ− is symmetric.

Even if Ω is not square, a unique Ω− can be found for Ω which satisfies the above four properties. This is called the Moore-Penrose generalized inverse. Note that a symmetric idempotent matrix is its own Moore-Penrose generalized inverse. For example, it is easy to verify that if Ω = PX , then Ω− = PX and that it satisfies the above four properties. Idempotent matrices have characteristic roots that are either zero or one. The number of non-zero characteristic roots is equal to the rank of this matrix. The characteristic roots of Ω−1 are the reciprocals of the characteristic roots of Ω, but the characteristic vectors of both matrices are the same. The determinant of a matrix is non-zero if and only if it has full rank. Therefore, if A is singular, then |A| = 0. Also, the determinant of a matrix is equal to the product of its characteristic roots. For two square matrices A and B, the determinant of the product is the product of the determinants |AB| = |A|·|B|. Therefore, the determinant of Ω−1 is the reciprocal of the determinant of Ω. This follows from the fact that |Ω||Ω−1 | = |ΩΩ−1 | = |I| = 1. This property is used in writing the likelihood function for Generalized Least Squares (GLS) estimation, see Chapter 9. The determinant of a triangular matrix


Appendix

173

is equal to the product of its diagonal elements. Of course, it immediately follows that the determinant of a diagonal matrix is the product of its diagonal elements. The constant in the regression corresponds to a vector of ones in the matrix of regressors X. This vector of ones is denoted by ιn where n is the dimension of this column vector. Note that ι′n ιn = n and ιn ι′n = Jn where Jn is a matrix of ones of dimension n × n. Note also that Jn is not idempotent, but ¯ ¯ J¯n = Jn /n is idempotent as can be easily verified. The rank(J¯n ) = tr(Jn ) = 1. Note also that In − Jn is idempotent with rank (n − 1). J¯n y has a typical element y¯ = ni=1 yi /n whereas (In − J¯n )y has a typical element (yi − y¯). So that J¯n is the averaging matrix, whereas premultiplying by (In − J¯n ) results in deviations from the mean. For two nonsingular matrices A and B (AB)−1 = B −1 A−1 Also, the transpose of a product of two conformable matrices, (AB)′ = B ′ A′ . In fact, for the product of three conformable matrices this becomes (ABC)′ = C ′ B ′ A′ . The transpose of the inverse is the inverse of the transpose, i.e., (A−1 )′ = (A′ )−1 . The inverse of a partitioned matrix

A=

A11 A21

A12 A22

is A−1 =

E −A−1 22 A21 E

−EA12 A−1 22 −1 −1 A22 + A−1 22 A21 EA12 A22

−1 where E = (A11 − A12 A−1 . Alternatively, it can be expressed as 22 A21 )

−1 −1 A11 + A11 A12 F A21 A−1 −A−1 A12 F −1 11 11 A = −F A21 A−1 F 11 −1 where F = (A22 − A21 A−1 . These formulas are used in partitioned regression models, see for 11 A12 ) example the Frisch-Waugh Lovell Theorem and the computation of the variance-covariance matrix of forecasts from a multiple regression in Chapter 7. An n × n symmetric matrix Ω has n distinct characteristic vectors c1 , . . . , cn . The corresponding n characteristic roots λ1 , . . . , λn may not be distinct but they are all real numbers. The number of nonzero characteristic roots of Ω is equal to the rank of Ω. The characteristic roots of a positive definite matrix are positive. The characteristic vectors of the symmetric matrix Ω are orthogonal to each other, i.e., c′i cj = 0 for i = j and can be made orthonormal with c′i ci = 1 for i = 1, 2, . . . , n. Hence, the matrix of characteristic vectors C = [c1 , c2 , . . . , cn ] is an orthogonal matrix, such that CC ′ = C ′ C = In with C ′ = C −1 . By definition Ωci = λi ci or ΩC = CΛ where Λ = diag[λi ]. Premultiplying the last vectors C diagonalizes equation by C ′ we get C ′ ΩC = C ′ CΛ = Λ. Therefore, the matrix of characteristic n the symmetric matrix Ω. Alternatively, we can write Ω = CΛC ′ = i=1 λi ci c′i which is the spectral decomposition of Ω. A real symmetric n × n matrix Ω is positive semi-definite if for every n × 1 non-negative vector y, we have y ′ Ωy ≥ 0. If y ′ Ωy is strictly positive for any non-zero y then Ω is said to be positive definite. A necessary and sufficient condition for Ω to be positive definite is that all the characteristic roots of Ω are positive. One important application is the comparison of efficiency of two unbiased estimators of a vector of parameters β. In this case, we subtract the variance-covariance matrix of the inefficient estimator from the more efficient one and show that the resulting difference yields a positive semi-definite matrix, see the Gauss-Markov Theorem in Chapter 7. If Ω is a symmetric and positive definite matrix, there exists a nonsingular matrix P such that Ω = P P ′ . In fact, using the spectral decomposition of Ω given above, one choice for P = CΛ1/2 so


174

Chapter 7: The General Linear Model: The Basics

that Ω = CΛC ′ = P P ′ . This is a useful result which we use in Chapter 9 to obtain Generalized Least Squares (GLS) as a least squares regression after transforming the original regression model by P −1 = (CΛ1/2 )−1 = Λ−1/2 C ′ . In fact, if u ∼ (0, σ 2 Ω), then P −1 u has zero mean and var(P −1 u) = P −1′ var(u)P −1′ = σ 2 P −1 ΩP −1′ = σ 2 P −1 P P ′ P −1′ = σ 2 In . From Chapter have seen that if u ∼ N (0, σ 2 In ), then ui /σ ∼ N (0, 1), so that u2i /σ 2 ∼ χ21 n2, we 2 ′ 2 and u u/σ = i=1 ui /σ 2 ∼ χ2n . Therefore, u′ (σ 2 In )−1 u ∼ χ2n . If u ∼ N (0, σ 2 Ω) where Ω is positive definite, then u∗ = P −1 u ∼ N (0, σ 2 In ) and u∗′ u∗ /σ 2 ∼ χ2n . But u∗′ u∗ = u′ P −1′ P −1 u = u′ Ω−1 u. Hence, u′ Ω−1 u/σ2 ∼ χ2n . This is used in Chapter 9. Note that the OLS residuals are denoted by e = P¯X u. If u ∼ N (0, σ 2 In ), then e has mean zero and var(e) = σ 2 P¯X In P¯X = σ 2 P¯X so that e ∼ N (0, σ 2 P¯X ). Our estimator of σ 2 in Chapter 7 is s2 = e′ e/(n−k) so that (n − k)s2 /σ 2 = e′ e/σ2 . The last term can also be written as u′ P¯X u/σ 2 . In order to find the distribution of this quadratic form in Normal variables, we use the following result stated as lemma 1 in Chapter 7. Lemma 1: For every symmetric idempotent matrix A of rank r, there exists an orthogonal matrix P such that P ′ AP = Jr where Jr is a diagonal matrix with the first r elements equal to one and the rest equal to zero. We use this lemma to show that the e′ e/σ 2 is a chi-squared with (n − k) degrees of freedom. To see this note that e′ e/σ2 = u′ P¯X u/σ2 and that P¯X is symmetric and idempotent of rank (n − k). Using the lemma there exists a matrix P such that P ′ P¯X P = Jn−k is a diagonal matrix with the first (n − k) elements on the diagonal equal to 1 and the last k elements equal to zero. An orthogonal matrix P is by definition a matrix whose inverse, is its own transpose, i.e., P ′ P = In . Let v = P ′ u then v has mean zero and var(v) = σ 2 P ′ P = σ 2 In so that v is N (0, σ 2 In ) and u = P v. Therefore, n−k 2 2 e′ e/σ 2 = u′ P¯X u/σ2 = v ′ P ′ P¯X P v/σ 2 = v ′ Jn−k v/σ 2 = i=1 vi /σ

But, the v’s are independent identically distributed N (0, σ 2 ), hence vi2 /σ 2 is the square of a standardized N (0, 1) random variable which is distributed as a χ21 . Moreover, the sum of independent χ2 random variables is a χ2 random variable with degrees of freedom equal to the sum of the respective degrees of freedom, see Chapter 2. Hence, e′ e/σ2 is distributed as χ2n−k . The beauty of the above result is that it applies to all quadratic forms u′ Au where A is symmetric and idempotent. In general, for u ∼ N (0, σ 2 I), a necessary and sufficient condition for u′ Au/σ 2 to be distributed χ2k is that A is idempotent of rank k, see Theorem 4.6 of Graybill (1961). Another useful theorem on quadratic forms in normal random variables is the following: If u ∼ N (0, σ 2 Ω), then u′ Au/σ 2 is χ2k if and only if AΩ is an idempotent matrix of rank k, see Theorem 4.8 of Graybill (1961). If u ∼ N (0, σ 2 I), the two positive semi-definite quadratic forms in normal random variables say u′ Au and u′ Bu are independent if and only if AB = 0, see Theorem 4.10 of Graybill (1961). A sufficient condition is that tr(AB) = 0, see Theorem 4.15 of Graybill (1961). This is used in Chapter 7 to construct F statistics to test hypotheses, see for example problem 11. For u ∼ N (0, σ 2 I), the quadratic form u′ Au is independent of the linear form Bu if BA = 0, see Theorem 4.17 of Graybill (1961). This is used in  , see problem 8. In general, if u ∼ N (0, Σ), then u′ Au Chapter 7 to prove the independence of s2 and β ols ′ and u Bu are independent if and only if AΣB = 0, see Theorem 4.21 of Graybill (1961). Many other useful matrix properties can be found. This is only a sample of them that will be implicitly or explicitly used in this book. The Kronecker product of two matrices say Σ⊗ In where Σ is m × m and In is the identity matrix of dimension n is defined as follows: ⎤ ⎡ σ 11 In . . . σ1m In ⎦ : : Σ ⊗ In = ⎣ σ m1 In . . . σ mm In

In other words, we place an In next to every element of Σ = [σ ij ]. The dimension of the resulting matrix is mn × mn. This is useful when we have a system of equations like Seemingly Unrelated Regressions in Chapter 10. In general, if A is m × n and B is p × q then A ⊗ B is mp × nq. Some properties of Kronecker


References

175

products include (A ⊗ B)′ = A′ ⊗ B ′ . If both A and B are square matrices of order m × m and p × p then (A ⊗ B)−1 = A−1 ⊗ B −1 , |A ⊗ B| = |A|m |B|p and tr(A ⊗ B) = tr(A)tr(B). Applying this result to Σ ⊗ In we get (Σ ⊗ In )−1 = Σ−1 ⊗ In

and

|Σ ⊗ In | = |Σ|m |In |n = |Σ|m

and tr(Σ ⊗ In ) = tr(Σ)tr(In ) = n tr(Σ). Some useful properties of matrix differentiation are the following: ∂x′ b =x ∂b

where x′ is 1 × k and b is k × 1.

Also ∂b′ Ab = (A + A′ ) ∂b

where A is k × k.

If A is symmetric, then ∂b′ Ab/∂b = 2Ab. These two properties will be used in Chapter 7 in deriving the least squares estimator.

References Bellman, R. (1970), Introduction to Matrix Analysis (McGraw Hill: New York). Searle, S.R. (1982), Matrix Algebra Useful for Statistics (John Wiley and Sons: New York).


CHAPTER 8

Regression Diagnostics and Specification Tests 8.1

Influential Observations1

Sources of influential observations include: (i) improperly recorded data, (ii) observational errors in the data, (iii) misspecification and (iv) outlying data points that are legitimate and contain valuable information which improve the efficiency of the estimation. It is constructive to isolate extreme points and to determine the extent to which the parameter estimates depend upon these desirable data. One should always run descriptive statistics on the data, see Chapter 2. This will often reveal outliers, skewness or multimodal distributions. Scatter diagrams should also be examined, but these diagnostics are only the first line of attack and are inadequate in detecting multivariate discrepant observations or the way each observation affects the estimated regression model. In regression analysis, we emphasize the importance of plotting the residuals against the explanatory variables or the predicted values y to identify patterns in these residuals that may indicate nonlinearity, heteroskedasticity, serial correlation, etc, see Chapter 3. In this section, we learn how to identify significantly large residuals and compute regression diagnostics that may identify influential observations. We study the extent to which the deletion of any observation affects the estimated coefficients, the standard errors, predicted values, residuals and test statistics. These represent the core of diagnostic tools in regression analysis. Accordingly, Belsley, Kuh and Welsch (1980, p.11) define an influential observation as “..one which, either individually or together with several other observations, has demonstrably larger impact on the calculated values of various estimates (coefficients, standard errors, t-values, etc.) than is the case for most of the other observations.” First, what is a significantly large residual? We have seen that the least squares residuals of y on X are given by e = (In − PX )u, see equation (7.7). y is n × 1 and X is n × k. If u ∼ IID(0, σ 2 In ), then e has zero mean and variance σ 2 (In − PX ). Therefore, the OLS residuals are correlated and heteroskedastic with var(ei ) = σ 2 (1 − hii ) where hii is the i-th diagonal element of the hat matrix H = PX , since y = Hy. The diagonal elements hii have the following properties: n  hii = nj=1 h2ij ≥ h2ii ≥ 0. i=1 hii = tr(PX ) = k and

The last property follows from the fact that PX is symmetric and idempotent. Therefore, h2ii − hii ≤ 0 or hii (hii − 1) ≤ 0. Hence, 0 ≤ hii ≤ 1, (see problem 1). hii is called the leverage of the i-th observation. For a simple regression with a constant,  hii = (1/n) + (x2i / ni=1 x2i ) ¯ hii can be interpreted as a measure of the distance between X values of where xi = Xi − X; the i-th observation and their mean over all n observations. A large hii indicates that the i-th observation is distant from the center of the observations. This means that the i-th observation with large hii (a function only of Xi values) exercises substantial leverage in determining the fitted value yi . Also, the larger hii , the smaller the variance of the residual ei . Since observations


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Chapter 8: Regression Diagnostics and Specification Tests

with high leverage tend to have smaller residuals, it may not be possible to detect them by an examination of the residuals alone. But, what is a large leverage? hii is large if it is more ¯ = 2k/n. Hence, hii ≥ 2k/n are considered outlying than twice the mean leverage value 2h observations with regards to X values. An alternative representation of hii is simply hii = d′i PX di = ||PX di ||2 = x′i (X ′ X)−1 xi where di denotes the i-th observation’s dummy variable, i.e., a vector of dimension n with 1 in the i-th position and 0 elsewhere. x′i is the i-th row of X and ||.|| denotes the Euclidian length. Note that d′i X = x′i . Let us standardize the i-th OLS residual by dividing it by an estimate of its variance. A standardized residual would then be:  (8.1) ei = ei /s 1 − hii where σ 2 is estimated by s2 , the MSE of the regression. This is an internal studentization of the residuals, see Cook and Weisberg (1982). Alternatively, one could use an estimate of σ 2 that is independent of ei . Defining s2(i) as the MSE from the regression computed without the i-th observation, it can be shown, see equation (8.18) below, that   n − k − e2i (n − k)s2 − e2i /(1 − hii ) (8.2) = s2 s2(i) = (n − k − 1) n−k−1

Under normality, s2(i) and ei are independent and the externally studentized residuals are defined by  e∗i = ei /s(i) 1 − hii ∼ tn−k−1 (8.3)

Thus, if the normality assumption holds, we can readily assess the significance of any single studentized residual. Of course, the e∗i will not be independent. Since this is a t-statistic, it is natural to think of e∗i as large if its value exceeds 2 in absolute value. Substituting (8.2) into (8.3) and comparing the result with (8.1), it is easy to show that e∗i is a monotonic transformation of ei e∗i

= ei



n−k−1 n − k − e2i

1 2

(8.4)

Cook and Wiesberg (1982) show that e∗i can be obtained as a t-statistic from the following augmented regression: y = Xβ ∗ + di ϕ + u

(8.5)

 = ei /(1 − hii ) and e∗i is the where di is the dummy variable for the i-th observation. In fact, ϕ t-statistic for testing that ϕ = 0. (see problem 4 and the proof given below). Hence, whether the i-th residual is large can be simply determined by the regression (8.5). A dummy variable for the i-th observation is included in the original regression and the t-statistic on this dummy tests whether this i-th residual is large. This is repeated for all observations i = 1, . . . , n. This can be generalized easily to testing for a group of significantly large residuals: y = Xβ ∗ + Dp ϕ∗ + u

(8.6)


8.1

Influential Observations

179

where Dp is an n × p matrix of dummy variables for the p-suspected observations. One can test ϕ∗ = 0 using the Chow test described in (4.17) as follows: F =

[Residual SS(no dummies) − Residual SS(Dp dummies used)]/p Residual SS(Dp dummies used)/(n − k − p)

(8.7)

This will be distributed as Fp,n−k−p under the null, see Gentleman and Wilk (1975). Let ep = Dp′ e,

then

E(ep ) = 0 and

var(ep ) = σ 2 Dp′ P¯X Dp

(8.8)

Then one can show, (see problem 5), that F =

[e′p (Dp′ P¯X Dp )−1 ep ]/p ∼ Fp ,n−k−p [(n − k)s2 − e′p (Dp′ P¯X Dp )−1 ep ]/(n − k − p)

(8.9)

Another refinement comes from estimating the regression without the i-th observation:  = [X ′ X(i) ]−1 X ′ y(i) β (i) (i) (i)

(8.10)

(A − a′ b)−1 = A−1 + A−1 a′ (I − bA−1 a′ )−1 bA−1

(8.11)

where the (i) subscript notation indicates that the i-th observation has been deleted. Using the updating formula

with A = (X ′ X) and a = b = x′i , one gets ′ [X(i) X(i) ]−1 = (X ′ X)−1 + (X ′ X)−1 xi x′i (X ′ X)−1 /(1 − hii )

(8.12)

Therefore −β  = (X ′ X)−1 xi ei /(1 − hii ) β (i)

(8.13)

Since the estimated coefficients are often of primary interest, (8.13) describes the change in the estimated regression coefficients that would occur if the i-th observation is deleted. Note that a high leverage observation with hii large will be influential in (8.13) only if the corresponding residual ei is not small. Therefore, high leverage implies a potentially influential observation, but whether this potential is actually realized depends on yi . Alternatively, one can obtain this result from the augmented regression given in (8.5). Note that Pdi = di (d′i di )−1 d′i = di d′i is an n × n matrix with 1 in the i-th diagonal position and 0 elsewhere. P¯di = In − Pdi , has the effect when post-multiplied by a vector y of deleting the i-th observation. Hence, premultiplying (8.5) by P¯di one gets       y(i) X(i) u(i) = β∗ + (8.14) P¯di y = 0 0 0 where the i-th observation is moved to the bottom of the data, without loss of generality. The last observation has no effect on the least squares estimate of β ∗ since both the dependent and ∗ = β  , and the i-th observation’s independent variables are zero. This regression will yield β (i) residual is clearly zero. By the Frisch-Waugh-Lovell Theorem given in section 7.3, the least squares estimates and the residuals from (8.14) are numerically identical to those from (8.5).


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Chapter 8: Regression Diagnostics and Specification Tests

∗ = β (i) in (8.5) and the i-th observation residual from (8.5) must be zero. This Therefore, β  , and the fitted values from this regression are given by y = X β  +di ϕ implies that ϕ  = yi −x′i β  (i) (i)  The difference in residuals is whereas those from the original regression (7.1) are given by X β. therefore  + di ϕ   − Xβ e − e(i) = X β (i)

(8.15)

ϕ  = ei /(1 − hii )

(8.16)

premultiplying (8.15) by P¯X and using the fact that P¯X X = 0, one gets P¯X (e − e(i) ) = P¯X di ϕ . But, P¯X e = e and P¯X e(i) = e(i) , hence P¯X di ϕ  = e−e(i) . Premultiplying both sides by d′i one gets  = ei since the i-th residual of e(i) from (8.5) is zero. By definition, d′i P¯X di = 1 − hii , d′i P¯X di ϕ therefore  −β  + (X ′ X)−1 X ′ di ϕ premultiplying (8.15) by (X ′ X)−1 X ′ one gets 0 = β  . This uses the fact (i) that both residuals are orthogonal to X. Rearranging terms and substituting ϕ  from (8.16), one gets −β  = (X ′ X)−1 xi ϕ  = (X ′ X)−1 xi ei /(1 − hii ) β (i)

as given in (8.13).  : Note that s2(i) given in (8.2) can now be written in terms of β (i) s2(i) =



t=i (yt

(i) )2 /(n − k − 1) − x′t β

(8.17)

upon substituting (8.13) in (8.17) we get

 e2i hit ei 2 − e + = (n − k − t t=1 1 − hii (1 − hii )2 n 2ei n e2i e2i 2 = (n − k)s2 + t=1 hit − t=1 et hit + 2 1 − hi (1 − hii ) (1 − hii )2 e2i = (n − k)s2 − (8.18) 1 − hii n which is (8.2). This uses the fact that He = 0 and H 2 = H. Hence, t=1 et hit = 0 and  n 2 =h . h ii it t=1  (the j-th component of β)  that results from the deletion To assess whether the change in β j j , σ 2 (X ′ X)−1 . This is of the i-th observation, is large or small, we scale by the variance of β jj denoted by  j − β  j(i) )/s(i) (X ′ X)−1 (8.19) DFBETAS ij = (β jj 1)s2(i)

n



Note that s(i) is used in order to make the denominator stochastically independent of the numerator in the Gaussian case. Absolute values of DFBETAS larger than 2 are considered √ influential. However, Belsley, Kuh, and Welsch (1980) suggest 2/ n as a size-adjusted cutoff. In fact, it would be most unusual for the removal of a single observation from a sample of 100


8.1

Influential Observations

181

or more to result in a change in any estimate by two or more standard errors. The size-adjusted cutoff tend to expose approximately the same proportion of potentially influential observations, regardless of sample size. The size-adjusted cutoff is particularly important for large data sets. In case of Normality, it can also be useful to look at the change in the t-statistics, as a means of assessing the sensitivity of the regression output to the deletion of the i-th observation:   β β j(i) j  − DFSTAT ij =  −1 ′ X )−1 s(i) (X(i) s (X ′ X)jj (i) jj

(8.20)

Another way to summarize coefficient changes and gain insight into forecasting effects when the i-th observation is deleted is to look at the change in fit, defined as −β  ] = hii ei /(1 − hii ) DFFIT i = yi − y(i) = x′i [β (i)

(8.21)

where the last equality is obtained from (8.13). √ We scale this measure by the variance of y(i) , i.e., σ hii , giving DFFITS i =



hii 1 − hii

1/2

e √i = s(i) 1 − hii



hii 1 − hii

1/2

e∗i

(8.22)

where σ has been estimated by s(i) and e∗i denotes the externally studentized residual given in (8.3). Values of DFFITS larger than 2 in absolute value are considered influential. A size adjusted cutoff for DFFITS suggested by Belsley, Kuh and Welsch (1980) is 2 k/n. In (8.3), the studentized residual e∗i was interpreted as a t-statistic that tests for the significance of the coefficient ϕ of di , the dummy variable which takes the value 1 for the i-th observation and 0 otherwise, in the regression of y on X and di . This can now be easily proved as follows: Consider the Chow test for the significance of ϕ. The RRSS = (n − k)s2 , the URSS = (n − k − 1)s2(i) and the Chow F -test described in (4.17) becomes F1,n−k−1 =

[(n − k)s2 − (n − k − 1)s2(i) ]/1 (n − k − 1)s2(i) /(n − k − 1)

=

e2i s2(i) (1 − hii )

(8.23)

The square root of (8.23) is e∗i ∼ tn−k−1 . These studentized residuals provide a better way to examine the information in the residuals, but they do not tell the whole story, since some of the most influential data points can have small e∗i (and very small ei ). One overall measure of the impact of the i-th observation on the estimated regression coefficients is Cook’s (1977) distance measure Di2 . Recall, that the confidence region for all k  − β)′ X ′ X(β  − β)/ks2 ∼ F (k, n − k). Cook’s (1977) distance mearegression coefficients is (β 2 sure Di uses the same structure for measuring the combined impact of the differences in the estimated regression coefficients when the i-th observation is deleted: −β  )′ X ′ X(β −β  )/ks2 Di2 (s) = (β (i) (i)

(8.24)

Even though Di2 (s) does not follow the above F -distribution, Cook suggests computing the percentile value from this F -distribution and declaring an influential observation if this percentile  and β  will be large, implying that the i-th value ≥ 50%. In this case, the distance between β (i)


182

Chapter 8: Regression Diagnostics and SpeciďŹ cation Tests

observation has a substantial inďŹ&#x201A;uence on the ďŹ t of the regression. Cookâ&#x20AC;&#x2122;s distance measure can be equivalently computed as:   hii e2i 2 (8.25) Di (s) = 2 ks (1 â&#x2C6;&#x2019; hii )2 Di2 (s) depends on ei and hii ; the larger ei or hii the larger is Di2 (s). Note the relationship between Cookâ&#x20AC;&#x2122;s Di2 (s) and Belsley, Kuh, and Welsch (1980) DFFITS i (Ď&#x192;) in (8.22), i.e.,  â&#x2C6;&#x161;  )/(Ď&#x192; hii ) DFFITS i (Ď&#x192;) = kDi (Ď&#x192;) = ( yi â&#x2C6;&#x2019; xâ&#x20AC;˛i β (i)

 Belsley, Kuh, and Welsch (1980) suggest nominating DFFITS based on s(i) exceeding 2 k/n â&#x2C6;&#x161; for special attention. Cookâ&#x20AC;&#x2122;s 50 percentile recommendation is equivalent to DFFITS > k, which is more conservative, see Velleman and Welsch (1981). Next, we study the inďŹ&#x201A;uence of the i-th observation deletion on the covariance matrix of the regression coeďŹ&#x192;cients. One can compare the two covariance matrices using the ratio of their determinants: ! â&#x20AC;˛ X ]â&#x2C6;&#x2019;1 â&#x20AC;˛ X ]â&#x2C6;&#x2019;1 ) det[X(i) s2k det(s2(i) [X(i) (i) (i) (i) (8.26) = 2k COVRATIO i = det(s2 [X â&#x20AC;˛ X]â&#x2C6;&#x2019;1 ) s det[X â&#x20AC;˛ X]â&#x2C6;&#x2019;1 Using the fact that â&#x20AC;˛ X(i) ] = (1 â&#x2C6;&#x2019; hii )det[X â&#x20AC;˛ X] det[X(i)

(8.27)

see problem 8, one obtains COVRATIO i =

s2(i) s2

!k

Ă&#x2014;

1 = 1 â&#x2C6;&#x2019; hii nâ&#x2C6;&#x2019;kâ&#x2C6;&#x2019;1 nâ&#x2C6;&#x2019;k

1 +

eâ&#x2C6;&#x2014;2 i nâ&#x2C6;&#x2019;k

k

(8.28) (1 â&#x2C6;&#x2019; hii )

where the last equality follows from (8.18) and the deďŹ nition of eâ&#x2C6;&#x2014;i in (8.3). Values of COVRATIO not near unity identify possible inďŹ&#x201A;uential observations and warrant further investigation. Belsley, Kuh and Welsch (1980) suggest investigating points with |COVRATIO â&#x2C6;&#x2019;1| near to or larger than 3k/n. The COVRATIO depends upon both hii and eâ&#x2C6;&#x2014;2 i . In fact, from (8.28), COVRATIO is large when hii is large and small when eâ&#x2C6;&#x2014;i is large. The two factors can offset each other, that is why it is important to look at hii and eâ&#x2C6;&#x2014;i separately as well as in combination as in COVRATIO. Finally, one can look at how the variance of yi changes when an observation is deleted. var( yi ) = s2 hii

and the ratio is

and

 ) = s2 (hii /(1 â&#x2C6;&#x2019; hii )) var( y(i) ) = var(xâ&#x20AC;˛i β (i) (i)

FVARATIO i = s2(i) /s2 (1 â&#x2C6;&#x2019; hii )

(8.29)

This expression is similar to COVRATIO except that [s2(i) /s2 ] is not raised to the k-th power. As a diagnostic measure it will exhibit the same patterns of behavior with respect to different conďŹ gurations of hii and the studentized residual as described for COVRATIO.


8.1

Influential Observations

183

Table 8.1 Cigarette Regression Dependent Variable: LNC Analysis of Variance Source Model Error C Total Root MSE Dep Mean C.V.

DF 2 43 45 0.16343 4.84784 3.37125

Sum of Squares 0.50098 1.14854 1.64953

Mean Square 0.25049 0.02671

R-square Adj R-sq

0.3037 0.2713

F Value 9.378

Prob>F 0.0004

T for H0: Parameter=0 4.730 –4.123 0.876

Prob > |T| 0.0001 0.0002 0.3858

Parameter Estimates

Variable INTERCEP LNP LNY

DF 1 1 1

Parameter Estimate 4.299662 –1.338335 0.172386

Standard Error 0.90892571 0.32460147 0.19675440

Example 1: For the cigarette data given in Table 3.2, Table 8.1 gives the SAS least squares regression for logC on logP and logY . logC = 4.30 − 1.34 logP + 0.172 logY + residuals (0.909) (0.325) (0.197) ¯ 2 = 0.271. Table 8.2 gives the data The standard error of the regression is s = 0.16343 and R along with the predicted values of logC, the least squares residuals e, the internal studentized residuals e given in (8.1), the externally studentized residuals e∗ given in (8.3), the Cook statistic given in (8.25), the leverage of each observation h, the DFFITS given in (8.22) and the COVRATIO given in (8.28). Using the leverage column, one can identify four potential observations with high leverage, i.e., ¯ = 2k/n = 6/46 = 0.13043. These are the observations belonging to the following greater than 2h states: Connecticut (CT), Kentucky (KY), New Hampshire (NH) and New Jersey (NJ) with leverage 0.13535, 0.19775, 0.13081 and 0.13945, respectively. Note that the corresponding OLS residuals are −0.078, 0.234, 0.160 and −0.059, which are not necessarily large. The internally studentized residuals are computed using equation (8.1). For KY this gives 0.23428 eKY √ = 1.6005 = eKY = √ s 1 − hKY 0.16343 1 − 0.19775

From Table 8.2, two observations with a high internally studentized residuals are those belonging to Arkansas (AR) and Utah (UT) with values of 2.102 and −2.679 respectively, both larger than 2 in absolute value. The externally studentized residuals are computed from (8.3). For KY, we first compute s2(KY ) , the MSE from the regression computed without the KY observation. From (8.2), this is


184

Chapter 8: Regression Diagnostics and Specification Tests

given by s2(KY ) = =

(n − k)s2 − e2KY /(1 − hKY ) (n − k − 1) (46 − 3)(0.16343)2 − (0.23428)2 /(1 − 0.19775) = 0.025716 (46 − 3 − 1)

From (8.3) we get e∗(KY ) =

0.23428 e √ √KY = 1.6311 = s(KY ) 1 − hKY 0.16036 1 − 0.19775

This externally studentized residual is distributed as a t-statistic with 42 degrees of freedom. However, e∗KY does not exceed 2 in absolute value. Again, e∗AR and e∗U T are 2.193 and −2.901 both larger than 2 in absolute value. From (8.13), the change in the regression coefficients due to the omission of the KY observation is given by ′ −1 −β  β (KY ) = (X X) xKY eKY /(1 − hKY )

Using the fact that

(X ′ X)−1

⎤ 30.929816904 4.8110214655 −6.679318415 = ⎣ 4.81102114655 3.9447686638 −1.177208398 ⎦ −6.679318415 −1.177208398 1.4493372835

and x′KY = (1, −0.03260, 4.64937) with eKY = 0.23428 and hKY = 0.19775 one gets ′ −β  (β (KY ) ) = (−0.082249, −0.230954, 0.028492)

In order to assess whether this change is large or small, we compute DFBETAS given in (8.19). For the KY observation, these are given by DFBETAS KY,1 =

 −β  β −0.082449 1 1(KY )  √ = = −0.09222 0.16036 30.9298169 s(KY ) (X ′ X)−1 11

Similarly, DFBETAS KY,2 = −0.7251 and DFBETAS KY,3 = 0.14758. These are not larger than √ √ 2 in absolute value. However, DFBETAS KY,2 is larger than 2/ n = 2/ 46 = 0.2949 in absolute value. This is the size-adjusted cutoff recommended by Belsley, Kuh and Welsch (1980) for large n. The change in the fit due to the omission of the KY observation is given by (8.21). In fact, −β  DFFIT KY = yKY − y(KY ) = x′KY [β (KY ) ] ⎛ ⎞ = (1, −0.03260, 4.64937) ⎜ −0.082249 ⎟ = 0.05775 ⎝ −0.230954 ⎠ −0.028492

or simply

DFFIT KY =

(0.19775)(0.23428) hKY eKY = = 0.05775 (1 − hKY ) 1 − 0.19775


8.2

Scaling it by the variance of y(KY ) we get from (8.22) DFFITS KY =



hKY 1 − hKY

1/2

e∗KY =



0.19775 1 − 0.19775

1/2

Recursive Residuals

185

(1.6311) = 0.8098

This larger than 2 in absolute value, but it is larger than the size-adjusted cutoff of  is not 2 k/n = 2 3/46 = 0.511. Note also that both DFFITS AR = 0.667 and DFFITS U T = −0.888 are larger than 0.511 in absolute value. Cook’s distance measure is given in (8.25) and for KY can be computed as      e2KY hKY (0.23428)2 0.19775 2 DKY (s) = = = 0.21046 ks2 (1 − hKY )2 3(0.16343)2 (1 − 0.19775)2 2 (s) = 0.13623 and D 2 (s) = 0.22399, The other two large Cook’s distance measures are DRAR UT respectively. COVRATIO omitting the KY observation can be computed from (8.28) as !k     s2(KY ) 1 1 0.025716 3 COVRATIO KY = = 1.1125 = s2 1 − hKY (0.16343)2 (1 − 0.019775)

which means that COVRATIO KY − 1/ = 0.1125 is less than 3k/n = 9/46 = 0.1956. Finally, FVARATIO omitting the KY observation can be computed from (8.29) as FVARATIO KY =

s2(KY ) s2 (1 − hKY )

=

0.025716 = 1.2001 (0.16343)2 (1 − 0.19775)

By several diagnostic measures, AR, KY and UT are influential observations that deserve special attention. The first two states are characterized with large sales of cigarettes. KY is a producer state with a very low price on cigarettes, while UT is a low consumption state due to its high percentage of Mormon population (a religion that forbids smoking). Table 8.3 gives the predicted consumption along with the 95% confidence band, the OLS residuals, and the internalized student residuals, Cook’s D-statistic and a plot of these residuals. This last plot highlights the fact that AR, UT and KY have large studentized residuals.

8.2

Recursive Residuals

In Section 8.1, we showed that the least squares residuals are heteroskedastic with non-zero covariances, even when the true disturbances have a scalar covariance matrix. This section studies recursive residuals which are a set of linear unbiased residuals with a scalar covariance matrix. They are independent and identically distributed when the true disturbances themselves are independent and identically distributed.2 These residuals are natural in time-series regressions and can be constructed as follows:  = (X ′ Xt )−1 X ′ Yt where Xt denotes 1. Choose the first t ≥ k observations and compute β t t t the t × k matrix of t observations on k variables and Yt′ = (y1 , . . . , yt ). The recursive residuals are basically standardized one-step ahead forecast residuals:   )/ 1 + x′ (X ′ Xt )−1 xt+1 wt+1 = (yt+1 − x′t+1 β (8.30) t t t+1


186

Chapter 8: Regression Diagnostics and Specification Tests

Table 8.2 Diagnostic Statistics for the Cigarettes Example STATE

LNC

LNP

LNY

e

e⋆

1

AL

4.96213

0.20487

4.64039

4.8254

0.1367

0.857

0.8546

0.012

0.0480

0.1919

2

AZ

4.66312

0.16640

4.68389

4.8844

–0.2213

–1.376

–1.3906

0.021

0.0315

–0.2508

0.9681

3

AR

5.10709

0.23406

4.59435

4.7784

0.3287

2.102

2.1932

0.136

0.0847

0.6670

0.8469

4

CA

4.50449

0.36399

4.88147

4.6540

–0.1495

–0.963

–0.9623

0.033

0.0975

–0.3164

1.1138

5

CT

4.66983

0.32149

5.09472

4.7477

–0.0778

–0.512

–0.5077

0.014

0.1354

–0.2009

1.2186

6

DE

5.04705

0.21929

4.87087

4.8458

0.2012

1.252

1.2602

0.018

0.0326

0.2313

0.9924

7

DC

4.65637

0.28946

5.05960

4.7845

–0.1281

–0.831

–0.8280

0.029

0.1104

–0.2917

1.1491

8

FL

4.80081

0.28733

4.81155

4.7446

0.0562

0.352

0.3482

0.002

0.0431

0.0739

1.1118 1.1142

OBS

PREDICTED

Cook’s D

Leverage

DFFITS COVRATIO 1.0704

9

GA

4.97974

0.12826

4.73299

4.9439

0.0358

0.224

0.2213

0.001

0.0402

0.0453

10

ID

4.74902

0.17541

4.64307

4.8653

–0.1163

–0.727

–0.7226

0.008

0.0413

–0.1500

1.0787

11

IL

4.81445

0.24806

4.90387

4.8130

0.0014

0.009

0.0087

0.000

0.0399

0.0018

1.1178

12

IN

5.11129

0.08992

4.72916

4.9946

0.1167

0.739

0.7347

0.013

0.0650

0.1936

1.1046

13

IA

4.80857

0.24081

4.74211

4.7949

0.0137

0.085

0.0843

0.000

0.0310

0.0151

1.1070

14

KS

4.79263

0.21642

4.79613

4.8368

–0.0442

–0.273

–0.2704

0.001

0.0223

–0.0408

1.0919

15

KY

5.37906

–0.03260

4.64937

5.1448

0.2343

1.600

1.6311

0.210

0.1977

0.8098

1.1126

16

LA

4.98602

0.23856

4.61461

4.7759

0.2101

1.338

1.3504

0.049

0.0761

0.3875

1.0224

17

ME

4.98722

0.29106

4.75501

4.7298

0.2574

1.620

1.6527

0.051

0.0553

0.4000

0.9403

18

MD

4.77751

0.12575

4.94692

4.9841

–0.2066

–1.349

–1.3624

0.084

0.1216

–0.5070

1.0731

19

MA

4.73877

0.22613

4.99998

4.8590

–0.1202

–0.769

–0.7653

0.018

0.0856

–0.2341

1.1258

20

MI

4.94744

0.23067

4.80620

4.8195

0.1280

0.792

0.7890

0.005

0.0238

0.1232

1.0518

21

MN

4.69589

0.34297

4.81207

4.6702

0.0257

0.165

0.1627

0.001

0.0864

0.0500

1.1724

22

MS

4.93990

0.13638

4.52938

4.8979

0.0420

0.269

0.2660

0.002

0.0883

0.0828

1.1712

23

MO

5.06430

0.08731

4.78189

5.0071

0.0572

0.364

0.3607

0.004

0.0787

0.1054

1.1541

24

MT

4.73313

0.15303

4.70417

4.9058

–0.1727

–1.073

–1.0753

0.012

0.0312

–0.1928

1.0210

25

NE

4.77558

0.18907

4.79671

4.8735

–0.0979

–0.607

–0.6021

0.003

0.0243

–0.0950

1.0719

26

NV

4.96642

0.32304

4.83816

4.7014

0.2651

1.677

1.7143

0.065

0.0646

0.4504

0.9366

27

NH

5.10990

0.15852

5.00319

4.9500

0.1599

1.050

1.0508

0.055

0.1308

0.4076

1.1422

28

NJ

4.70633

0.30901

5.10268

4.7657

–0.0594

–0.392

–0.3879

0.008

0.1394

–0.1562

1.2337

29

NM

4.58107

0.16458

4.58202

4.8693

–0.2882

–1.823

–1.8752

0.076

0.0639

–0.4901

0.9007

30

NY

4.66496

0.34701

4.96075

4.6904

–0.0254

–0.163

–0.1613

0.001

0.0888

–0.0503

1.1755

31

ND

4.58237

0.18197

4.69163

4.8649

–0.2825

–1.755

–1.7999

0.031

0.0295

–0.3136

0.8848

32

OH

4.97952

0.12889

4.75875

4.9475

0.0320

0.200

0.1979

0.001

0.0423

0.0416

1.1174

33

OK

4.72720

0.19554

4.62730

4.8356

–0.1084

–0.681

–0.6766

0.008

0.0505

–0.1560

1.0940

34

PA

4.80363

0.22784

4.83516

4.8282

–0.0246

–0.153

–0.1509

0.000

0.0257

–0.0245

1.0997

35

RI

4.84693

0.30324

4.84670

4.7293

0.1176

0.738

0.7344

0.010

0.0504

0.1692

1.0876

36

SC

5.07801

0.07944

4.62549

4.9907

0.0873

0.555

0.5501

0.008

0.0725

0.1538

1.1324

37

SD

4.81545

0.13139

4.67747

4.9301

–0.1147

–0.716

–0.7122

0.007

0.0402

–0.1458

1.0786

38

TN

5.04939

0.15547

4.72525

4.9062

0.1432

0.890

0.8874

0.008

0.0294

0.1543

1.0457

39

TX

4.65398

0.28196

4.73437

4.7384

–0.0845

–0.532

–0.5271

0.005

0.0546

–0.1267

1.1129

40

UT

4.40859

0.19260

4.55586

4.8273

–0.4187

–2.679

–2.9008

0.224

0.0856

–0.8876

0.6786

41

VT

5.08799

0.18018

4.77578

4.8818

0.2062

1.277

1.2869

0.014

0.0243

0.2031

0.9794

42

VA

4.93065

0.11818

4.85490

4.9784

–0.0478

–0.304

–0.3010

0.003

0.0773

–0.0871

1.1556

43

WA

4.66134

0.35053

4.85645

4.6677

–0.0064

–0.041

–0.0404

0.000

0.0866

–0.0124

1.1747

44

WV

4.82454

0.12008

4.56859

4.9265

–0.1020

–0.647

–0.6429

0.011

0.0709

–0.1777

1.1216

45

WI

4.83026

0.22954

4.75826

4.8127

0.0175

0.109

0.1075

0.000

0.0254

0.0174

1.1002

46

WY

5.00087

0.10029

4.71169

4.9777

0.0232

0.146

0.1444

0.000

0.0555

0.0350

1.1345


Table 8.3 Regression of Real Per-Capita Consumption of Cigarettes Dep Obs

4.8254 4.8844 4.7784 4.6540 4.7477 4.8458 4.7845 4.7446 4.9439 4.8653 4.8130 4.9946 4.7949 4.8368 5.1448 4.7759 4.7298 4.9841 4.8590 4.8195 4.6702 4.8979 5.0071 4.9058 4.8735 4.7014 4.9500 4.7657 4.8693 4.6904 4.8649 4.9475 4.8356 4.8282 4.7293 4.9907 4.9301 4.9062 4.7384 4.8273 4.8818 4.9784 4.6677 4.9265 4.8127 4.9777

0.036 0.029 0.048 0.051 0.060 0.030 0.054 0.034 0.033 0.033 0.033 0.042 0.029 0.024 0.073 0.045 0.038 0.057 0.048 0.025 0.048 0.049 0.046 0.029 0.025 0.042 0.059 0.061 0.041 0.049 0.028 0.034 0.037 0.026 0.037 0.044 0.033 0.028 0.038 0.048 0.025 0.045 0.048 0.044 0.026 0.039 0 1.1485 1.3406

Lower95% Mean 4.7532 4.8259 4.6825 4.5511 4.6264 4.7863 4.6750 4.6761 4.8778 4.7983 4.7472 4.9106 4.7368 4.7876 4.9982 4.6850 4.6523 4.8692 4.7625 4.7686 4.5733 4.8000 4.9147 4.8476 4.8221 4.6176 4.8308 4.6427 4.7859 4.5922 4.8083 4.8797 4.7616 4.7754 4.6553 4.9020 4.8640 4.8497 4.6614 4.7308 4.8304 4.8868 4.5708 4.8387 4.7602 4.9000

Upper95% Mean 4.8976 4.9429 4.8743 4.7570 4.8689 4.9053 4.8940 4.8130 5.0100 4.9323 4.8789 5.0786 4.8529 4.8860 5.2913 4.8668 4.8074 5.0991 4.9554 4.8703 4.7671 4.9959 5.0996 4.9640 4.9249 4.7851 5.0692 4.8888 4.9526 4.7886 4.9215 5.0153 4.9097 4.8811 4.8033 5.0795 4.9963 4.9626 4.8155 4.9237 4.9332 5.0701 4.7647 5.0143 4.8653 5.0553

Lower95% Predict

Upper95% Predict

Std Err Residual

Student Residual

Residual

4.4880 4.5497 4.4351 4.3087 4.3965 4.5109 4.4372 4.4079 4.6078 4.5290 4.4769 4.6544 4.4602 4.5036 4.7841 4.4340 4.3912 4.6351 4.5155 4.4860 4.3267 4.5541 4.6648 4.5711 4.5399 4.3613 4.5995 4.4139 4.5293 4.3465 4.5305 4.6110 4.4978 4.4944 4.3915 4.6494 4.5940 4.5718 4.4000 4.4839 4.5482 4.6363 4.3242 4.5854 4.4790 4.6391

5.1628 5.2191 5.1217 4.9993 5.0989 5.1808 5.1318 5.0812 5.2801 5.2016 5.1491 5.3347 5.1295 5.1701 5.5055 5.1178 5.0684 5.3332 5.2024 5.1530 5.0137 5.2418 5.3495 5.2405 5.2071 5.0414 5.3005 5.1176 5.2092 5.0343 5.1993 5.2840 5.1735 5.1621 5.0671 5.3320 5.2663 5.2406 5.0769 5.1707 5.2154 5.3205 5.0113 5.2676 5.1465 5.3163

0.1367 –0.2213 0.3287 –0.1495 –0.0778 0.2012 –0.1281 0.0562 0.0358 –0.1163 0.00142 0.1167 0.0137 –0.0442 0.2343 0.2101 0.2574 –0.2066 –0.1202 0.1280 0.0257 0.0420 0.0572 –0.1727 –0.0979 0.2651 0.1599 –0.0594 –0.2882 –0.0254 –0.2825 0.0320 –0.1084 –0.0246 0.1176 0.0873 –0.1147 0.1432 –0.0845 –0.4187 0.2062 –0.0478 –0.00638 –0.1020 0.0175 0.0232

0.159 0.161 0.156 0.155 0.152 0.161 0.154 0.160 0.160 0.160 0.160 0.158 0.161 0.162 0.146 0.157 0.159 0.153 0.156 0.161 0.156 0.156 0.157 0.161 0.161 0.158 0.152 0.152 0.158 0.156 0.161 0.160 0.159 0.161 0.159 0.157 0.160 0.161 0.159 0.156 0.161 0.157 0.156 0.158 0.161 0.159

0.857 –1.376 2.102 –0.963 –0.512 1.252 –0.831 0.352 0.224 –0.727 0.009 0.739 0.085 –0.273 1.600 1.338 1.620 –1.349 –0.769 0.792 0.165 0.269 0.364 –1.073 –0.607 1.677 1.050 –0.392 –1.823 –0.163 –1.755 0.200 –0.681 –0.153 0.738 0.555 –0.716 0.890 –0.532 –2.679 1.277 –0.304 –0.041 –0.647 0.109 0.146

–2 –1 0

1 2 

     

 

     

             



Cook’s D 0.012 0.021 0.136 0.033 0.014 0.018 0.029 0.002 0.001 0.008 0.000 0.013 0.000 0.001 0.210 0.049 0.051 0.084 0.018 0.005 0.001 0.002 0.004 0.012 0.003 0.065 0.055 0.008 0.076 0.001 0.031 0.001 0.008 0.000 0.010 0.008 0.007 0.008 0.005 0.224 0.014 0.003 0.000 0.011 0.000 0.000

187

Sum of Residuals Sum of Squared Residuals Predicted Resid SS (Press)

Std Err Predict

Recursive Residuals

4.9621 4.6631 5.1071 4.5045 4.6698 5.0471 4.6564 4.8008 4.9797 4.7490 4.8145 5.1113 4.8086 4.7926 5.3791 4.9860 4.9872 4.7775 4.7388 4.9474 4.6959 4.9399 5.0643 4.7331 4.7756 4.9664 5.1099 4.7063 4.5811 4.6650 4.5824 4.9795 4.7272 4.8036 4.8469 5.0780 4.8155 5.0494 4.6540 4.4086 5.0880 4.9307 4.6613 4.8245 4.8303 5.0009

Predict Value

8.2

1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 24 25 26 27 28 29 30 31 32 33 34 35 36 37 38 39 40 41 42 43 44 45 46

Var LNC


188

Chapter 8: Regression Diagnostics and Specification Tests

 t+1 = (X ′ Xt+1 )−1 X ′ Yt+1 . 2. Add the (t + 1)-th observation to the data and obtain β t+1 t+1 Compute wt+2 . 3. Repeat step 2, adding one observation at a time. In time-series regressions, one usually starts with the first k-observations and obtain (T − k) forward recursive residuals. These recursive residuals can be computed using the updating formula given in (8.11) with A = (Xt′ Xt ) and a = −b = x′t+1 . Therefore, ′ (Xt+1 Xt+1 )−1 = (Xt′ Xt )−1 −(Xt′ Xt )−1 xt+1 x′t+1 (Xt′ Xt )−1 /[1+x′t+1 (Xt′ Xt )−1 xt+1 ] (8.31)

and only (Xt′ Xt )−1 have to be computed. Also, ′ −1 ′    β t+1 = β t + (Xt Xt ) xt+1 (yt+1 − xt+1 β t )/ft+1

(8.32)

where ft+1 = 1 + x′t+1 (Xt′ Xt )−1 xt+1 , see problem 13. Alternatively, one can compute these residuals by regressing Yt+1 on Xt+1 and dt+1 where dt+1 = 1 for the (t + 1)-th observation, and zero otherwise, see equation (8.5). The estimated coefficient of dt+1 is the numerator of wt+1 . The standard error of this estimate is st+1 times the denominator of wt+1 , where st+1 is the standard error of this regression. Hence, wt+1 can be retrieved as st+1 multiplied by the t-statistic corresponding to dt+1 . This computation has to be performed sequentially, in each case generating the corresponding recursive residual. This may be computationally inefficient, but it is simple to generate using regression packages. It is obvious from (8.30) that if ut ∼ IIN(0, σ 2 ), then wt+1 has zero mean and var(wt+1 ) = σ 2 . Furthermore, wt+1 is linear in the y’s. Therefore, it is normally distributed. It remains to show that the recursive residuals are independent. Given normality, it is sufficient to show that cov(wt+1 , ws+1 ) = 0

for

t = s; t, s = k, . . . , T − 1

(8.33)

This is left as an exercise for the reader, see problem 13. Alternatively, one can express the T − k vector of recursive residuals as w = Cy where C is of dimension (T − k) × T as follows: ⎤ ⎡ x′ (X ′ X )−1 X ′ k k √1 0....0 − k+1 √k fk+1 fk+1 ⎥ ⎢ ⎥ ⎢ .. .. ⎥ ⎢ . . ⎥ ⎢ ′ ′ ⎥ ⎢ x′t (Xt−1 Xt−1 )−1 Xt−1 (8.34) C=⎢ − √ √1 0....0 ⎥ ft ft ⎥ ⎢ ⎥ ⎢ .. .. ⎥ ⎢ . . ⎦ ⎣ x′T (XT′ −1 XT −1 )−1 XT′ −1 √ √1 − f f T

T

Problem 14 asks the reader to verify that w = Cy, using (8.30). Also, that the matrix C satisfies the following properties: (i) CX = 0

(ii) CC ′ = IT −k

(iii) C ′ C = P¯X

(8.35)

This means that the recursive residuals w are (LUS) linear in y, unbiased with mean zero and have a scalar variance-covariance matrix: var(w) = CE(uu′ )C ′ = σ 2 IT −k . Property (iii) also


8.2

Recursive Residuals

189

means that w′ w = y ′ C ′ Cy = y ′ P¯X y = e′ e. This means that the sum of squares of (T − k) recursive residuals is equal to the sum of squares of T least squares residuals. One can also show from (8.32) that 2 RSSt+1 = RSSt + wt+1

for

t = k, . . . , T − 1

(8.36)

 t ), see problem 14. Note that for t = k; RSS = 0, since  t )′ (Yt − Xt β where RSSt = (Yt − Xt β with k observations one gets a perfect fit and zero residuals. Therefore RSST =

T

2 t=k+1 wt

=

T

2 t=1 et

(8.37)

Applications of Recursive Residuals Recursive residuals have been used in several important applications: (1) Harvey (1976) used these recursive residuals to give an alternative proof of the fact that Chow’s post-sample predictive test has an F -distribution. Recall, from Chapter 7, that when the second sample n2 had fewer than k observations, Chow’s test becomes F =

(e′ e − e′1 e1 )/n2 ∼ F (n2 , n1 − k) e′1 e1 /(n1 − k)

(8.38)

where e′ e = RSS from the total sample (n1 + n2 = T observations), and e′1 e1 = RSS from the first n1 observations. Recursive residuals can be computed for t = k + 1, . . . , n1 , and continued on for the extra n2 observations. From (8.36) we have e′ e = Therefore,

n1 +n2

2 t=k+1 wt

and

n1 +n2 2 wt /n2 F = n1 t=n1 +1 2 t=k+1 wt /(n1 − k)

e′1 e1 =

n1

2 t=k+1 wt

(8.39)

(8.40)

But the wt ’s are ∼IIN(0, σ 2 ) under the null, therefore the F -statistic in (8.38) is a ratio of two independent chi-squared variables, each divided by the appropriate degrees of freedom. Hence, F ∼ F (n2 , n1 − k) under the null, see Chapter 2. (2) Harvey and Phillips (1974) used recursive residuals to test the null hypothesis of homoskedasticity. If the alternative hypothesis is that σ 2i varies with Xj , the proposed test is as follows: 1) Order the data according to Xj and choose a base of at least k observations from among the central observations. 2) From the first m observations compute the vector of recursive residuals w1 using the base constructed in step 1. Also, compute the vector of recursive residuals w2 from the last m observations. The maximum m can be is (T − k)/2.


190

Chapter 8: Regression Diagnostics and Specification Tests

3) Under the null hypothesis, it follows that F = w2′ w2 /w1′ w1 ∼ Fm,m

(8.41)

Harvey and Phillips suggest setting m at approximately (n/3) provided n > 3k. This test has the advantage over the Goldfeld-Quandt test in that if one wanted to test whether σ 2i varies with some other variable Xs , one could simply regroup the existing recursive residuals according to low and high values of Xs and compute (8.41) afresh, whereas the Goldfeld-Quandt test would require the computation of two new regressions. (3) Phillips and Harvey (1974) suggest using the recursive residuals to test the null hypothesis of no serial correlation using a modified von Neuman ratio: T (wt − wt−1 )2 /(T − k − 1) (8.42) M V N R = t=k+2T 2 t=k+1 wt /(T − k)

This is the ratio of the mean-square successive difference to the variance. It is arithmetically closely related to the DW statistic, but given that w ∼ N (0, σ 2 IT −k ) one has an exact test available and no inconclusive regions. Phillips and Harvey (1974) provide tabulations of the significance points. If the sample size is large, a satisfactory approximation is obtained from a normal distribution with mean 2 and variance 4/(T − k).

(4) Harvey and Collier (1977) suggest a test for functional misspecification based on recursive residuals. This is based on the fact that w ∼ N (0, σ 2 IT −k ). Therefore, √ w/(s ¯ w / T − k) ∼ tT −k−1 (8.43)   where w ¯ = Tt=k+1 wt /(T − k) and s2w = Tt=k+1 (wt − w) ¯ 2 /(T − k − 1). Suppose that the true functional form relating y to a single explanatory variable X is concave (convex) and the data are ordered by X. A simple linear regression is estimated by regressing y on X. The recursive residuals would be expected to be mainly negative (positive) and the computed t-statistic will be large in absolute value. When there are multiple X’s, one could carry out this test based on any single explanatory variable. Since several specification errors might have a self-cancelling effect on the recursive residuals, this test is not likely to be very effective in multivariate situations. Recently, Wu (1993) suggested performing this test using the following augmented regression: y = Xβ + zγ + v

(8.44)

where z = C ′ ιT −k is one additional regressor with C defined in (8.34) and ιT −k denoting a vector of ones of dimension T − k. In fact, the F -statistic for testing H0 ; γ = 0 turns out to be the square of the Harvey and Collier (1977) t-statistic given in (8.43), see problem 15. Alternatively, a Sign test may be used to test the null hypothesis of no functional misspecification. Under the null hypothesis, the expected number of positive recursive residuals is equal to (T − k)/2. A critical region may therefore be constructed from the binomial distribution. However, Harvey and Collier (1977) suggest that the Sign test tends to lack power compared with the t-test described in (8.43). Nevertheless, it is very simple and it may be more robust to non-normality.


8.2

Recursive Residuals

191

(5) Brown, Durbin and Evans (1975) used recursive residuals to test for structural change over time. The null hypothesis is  β 1 = β 2 = .. = β T = β (8.45) H0 ; σ 21 = σ 22 = .. = σ 2T = σ 2 where β t is the vector of coefficients in period t and σ 2t is the disturbance variance for that period. The authors suggest a pair of tests. The first is the CUSUM test which computes  Wr = rt=k+1 wt /sw for r = k + 1, . . . , T (8.46)

where s2w is an estimate of the variance of the wt ’s, given below (8.43). Wr is a cumulative sum and should be plotted against r. Under the null, E(Wr ) = 0. But, if there is a structural break, Wr will tend to diverge from the horizontal line. The authors suggest checking whether Wr  √ T − k and cross a pair of straight lines (see Figure 8.1) which pass through the points k, ±a  √  T, ±3a T − k where a depends upon the chosen significance level α. For example, a = 0.850, 0.948, and 1.143 for α = 10%, 5%, and 1% levels, respectively. If the coefficients are not constant, there may be a tendency for a disproportionate number of recursive residuals to have the same sign and to push Wr across the boundary. The second test is the cumulative sum of squares (CUSUMSQ) which is based on plotting   Wr∗ = rt=k+1 wt2 / Tt=k+1 wt2 for t = k + 1, . . . , T (8.47)

against r. Under the null, E(Wr∗ ) = (r − k)/(T − k) which varies from 0 for r = k to 1 for r = T . The significance of the departure of Wr∗ from its expected value is assessed by whether Wr∗ crosses a pair of lines parallel to E(Wr∗ ) at a distance co above and below this line. Brown, Durbin and Evans (1975) provide values of c0 for various sample sizes T and levels of significance α. The CUSUM and CUSUMSQ should be regarded as data analytic techniques; i.e., the value of the plots lie in the information to be gained simply by inspecting them. The plots contain more information than can be summarized in a single test statistic. The significance lines constructed are, to paraphrase the authors, best regarded as ‘yardsticks’ against which to assess the observed plots rather than as formal tests of significance. See Brown et al. (1975) for various examples. Note that the CUSUM and CUSUMSQ are quite general tests for structural change in that they do not require a prior determination of where the structural break takes place. If this is known, the Chow-test will be more powerful. But, if this break is not known, the CUSUM and CUSUMSQ are more appropriate. Example 2: Table 8.4 reproduces the consumption-income data, over the period 1950-1993, taken from the Economic Report of the President. In addition, the recursive residuals are computed as in (8.30) and exhibited in column 4, starting with 1952 and ending in 1993. Column 5 gives the CUSUM given by Wr in (8.46) and this is plotted against r in Figure 8.2. The CUSUM crosses the upper 5% line in 1993, showing structural instability in the latter years. This was done using EViews. The post-sample predictive test for 1993, can be obtained from (8.38) by computing the RSS from 1950-1992 and comparing it with the RSS from 1950-1993. The observed F -statistic is 5.3895 which is distributed as F (1, 41). The same F -statistic is obtained from (8.40) using recursive residuals. In fact,  2 2 2 / 1992 F = w1993 t=1952 wt /41 = (339.300414) /875798.36 ÷ 41 = 5.3895


192

Chapter 8: Regression Diagnostics and Specification Tests

wr

3a T – k 2a T – k

a T –k 0

n

t

k

–a T – k

– 3a T – k

Figure 8.1 CUSUM Critical Values

The Phillips and Harvey (1974) modified von Neuman ratio can be computed from (8.42) and yields an MVNR of 0.4263. The asymptotic distribution of this statistic is N (2, 4/(44 − 2)) which yields a standardized N (0, 1) statistic z = (0.4263 − 2)/0.3086 = −5.105. This is highly significant and indicates the presence of positive serial correlation. The Harvey and Collier (1977) functional misspecification test, given in (8.43), yields w¯ = 66.585 and sw = 140.097 and an observed t-statistic of 0.0733. This is distributed as t41 under the null hypothesis and is not significant. 30 20 10 0 –10

–20 55

60

65 CUSUM

70

75

80

85

90

5% Significance

Figure 8.2 CUSUM Plot of Consumption-Income Data


8.2

Recursive Residuals

193

Table 8.4 Consumption-Income Example YEAR 1950 1951 1952 1953 1954 1955 1956 1957 1958 1959 1960 1961 1962 1963 1964 1965 1966 1967 1968 1969 1970 1971 1972 1973 1974 1975 1976 1977 1978 1979 1980 1981 1982 1983 1984 1985 1986 1987 1988 1989 1990 1991 1992 1993

Consumption 5820 5843 5917 6054 6099 6365 6440 6465 6449 6658 6698 6740 6931 7089 7384 7703 8005 8163 8506 8737 8842 9022 9425 9752 9602 9711 10121 10425 10744 10876 10746 10770 10782 11179 11617 12015 12336 12568 12903 13029 13093 12899 13110 13391

Income

Recursive residuals

Wr of CUSUM

6284 6390 6476 6640 6628 6879 7080 7114 7113 7256 7264 7382 7583 7718 8140 8508 8822 9114 9399 9606 9875 10111 10414 11013 10832 10906 11192 11406 11851 12039 12005 12156 12146 12349 13029 13258 13552 13545 13890 14005 14101 14003 14279 14341

N/A N/A 24.90068062 30.35482733 50.89329055 63.26038894 –49.80590703 –28.40431109 –31.52055902 53.19425584 67.69611375 –2.64655577 9.67914674 39.65882720 –40.12655671 –30.26075567 2.60563312 –78.94146710 27.18506579 64.36319479 –64.90671688 –71.64101261 70.09586688 –113.47532273 –85.63317118 –29.42763002 128.32845919 220.69313327 126.59174925 78.39424741 –25.95557363 –124.17868561 –90.84519334 127.83058064 –30.79462855 159.78087196 201.70712713 405.31056105 390.95384121 373.37091949 316.43123530 188.10968277 134.46128461 339.30041400

N/A N/A 0.17773916 0.39440960 0.75768203 1.20922986 0.85371909 0.65097129 0.42597994 0.80567648 1.28888618 1.26999527 1.33908428 1.62216596 1.33574566 1.11974669 1.13834551 0.57486733 0.76891226 1.22833184 0.76503264 0.25366456 0.75400351 –0.05597469 –0.66721774 –0.87726992 0.03872884 1.61401959 2.51762185 3.07719401 2.89192510 2.00554711 1.35710105 2.26954599 2.04973628 3.16024996 4.63001005 7.52308592 10.31368463 12.97877777 15.23743980 16.58015237 17.53992676 19.96182724


194

Chapter 8: Regression Diagnostics and Specification Tests

8.3

Specification Tests

Specification tests are an important part of model specification in econometrics. In this section, we only study a few of these diagnostic tests. For an excellent summary on this topic, see Wooldridge (2001). (1) Ramsey’s (1969) RESET (Regression Specification Error Test) Ramsey suggests testing the specification of the linear regression model yt = Xt′ β + ut by augmenting it with a set of regressors Zt so that the augmented model is yt = Xt′ β + Zt′ γ + ut

(8.48)

If the Zt ’s are available then the specification test would reduce to the F -test for H0 ; γ = 0. The crucial issue is the choice of Zt variables. This depends upon the true functional form under the alternative, which is usually unknown. However, this can be often well approximated by higher powers of the initial regressors, as in the case where the true form is quadratic or cubic. OLS . The popular Alternatively, one might approximate it with higher moments of yˆt = Xt′ β Ramsey RESET test is carried out as follows: (1) Regress yt on Xt and get yˆt . (2) Regress yt on Xt , yˆt2 , yˆt3 and yˆt4 and test that the coefficients of all the powers of yˆt are zero. This is an F3,T −k−3 under the null. Note that yˆt is not included among the regressors because it would be perfectly multicollinear with Xt .3 Different choices of Zt ’s may result in more powerful tests when H0 is not true. Thursby and Schmidt (1977) carried out an extensive Monte Carlo and concluded that the test based on Zt = [Xt2 , Xt3 , Xt4 ] seems to be generally the best choice. (2) Utts’ (1982) Rainbow Test The basic idea behind the Rainbow test is that even when the true relationship is nonlinear, a good linear fit can still be obtained over subsets of the sample. The test therefore rejects the null hypothesis of linearity whenever the overall fit is markedly inferior to the fit over a properly selected sub-sample of the data, see Figure 8.3. e be Let e′ e be the OLS residuals sum of squares from all available n observations and let e′  the OLS residual sum of squares from the middle half of the observations (T /2). Then  (e′ e − e′ e)/ T2  (8.49) is distributed as F T ,( T −k) under H0 F = 2 2 e′ e/ T − k 2

 Under H0 ; E(e′ e/T − k) = σ 2 = E e′ e/ T2 − k , while in general under HA ; E(e′ e/T − k) >

′ T E e e/ 2 − k > σ 2 . The RRSS is e′ e because all the observations are forced to fit the straight line, whereas the URSS is e′ e because only a part of the observations are forced to fit a straight line. The crucial issue of the Rainbow test is the proper choice of the subsample (the middle T /2 observations in case of one regressor). This affects the power of the test and


8.3

Specification Tests

195

y

. . . . . .. . . . . . . . . . . . . . . .  . .  . . . . .

X

Figure 8.3 The Rainbow Test

not the distribution of the test statistic under the null. Utts (1982) recommends points close ¯ since an incorrect linear fit will in general not be as far off there as it is in the outer to X, ¯ is measured by the magnitude of the corresponding diagonal elements region. Closeness to X of PX . Close points are those with low leverage hii , see section 8.1. The optimal size of the subset depends upon the alternative. Utts recommends about 1/2 of the data points in order to obtain some robustness to outliers. The F -test in (8.49) looks like a Chow test, but differs in the selection of the sub-sample. For example, using the post-sample predictive Chow test, the data are arranged according to time and the first T observations are selected. The Rainbow ¯ and selects the first T /2 of them. test arranges the data according to their distance from X

(3) Plosser, Schwert and White (1982) (PSW) Differencing Test The differencing test is a general test for misspecification (like Hausman’s (1978) test, which will be introduced in the simultaneous equation chapter) but for time-series data only. This test compares OLS and First Difference (FD) estimates of β. Let the differenced model be ˙ + u˙ y˙ = Xβ

(8.50)

where y˙ = Dy, X˙ = DX and u˙ = Du where ⎡

⎢ ⎢ D=⎢ ⎢ ⎣

1 −1 0 0 ... 0 1 −1 0 . . . 0 0 1 −1 . . . . . . . ... 0 0 0 0 ...

0 0 0 0 0 0 . . 1 −1

⎥ ⎥ ⎥ is the familiar (T − 1) × T differencing matrix. ⎥ ⎦

Wherever there is a constant in the regression, the first column of X becomes zero and is dropped. From (8.50), the FD estimator is given by  ˙ ′ ˙ −1 ˙ ′ β F D = (X X) X y˙

(8.51)


196

Chapter 8: Regression Diagnostics and Specification Tests

 F D ) = σ 2 (X˙ ′ X) ˙ −1 since var(u) ˙ −1 X˙ ′ DD′ X( ˙ X˙ ′ X) with var(β ˙ = σ 2 (DD)′ and ⎡ ⎤ 2 −1 0 . . . 0 0 ⎢ −1 2 −1 . . . 0 0 ⎥ ⎢ ⎥ DD′ = ⎢ : : : : : : ⎥ ⎢ ⎥ ⎣ 0 0 0 . . . 2 −1 ⎦ 0 0 0 . . . −1 2

The differencing test is based on   q = β F D − β OLS

 ) − V (β  with V ( q ) = σ 2 [V (β FD OLS )]

A consistent estimate of V ( q ) is ⎡ !−1 ! ˙ ′ DD′ X˙ ˙ ′ X˙ X X 2 V ( q) = σ  ⎣ T T

X˙ ′ X˙ T

!−1



X ′X T

(8.52)

−1

where σ 2 is a consistent estimate of σ 2 . Therefore,

⎤ ⎦

(8.53)

Δ = T q′ [V ( q )]−1 q ∼ χ2k under H0

(8.54)

yt = β 1 x1t + β 2 x2t + ut

(8.55)

where k is the number of slope parameters if V ( q ) is nonsingular. V ( q ) could be singular, in −  which case we use a generalized inverse V ( q ) of V ( q ) and in this case is distributed as χ2 with degrees of freedom equal to the rank(V ( q )). This is a special case of the general Hausman (1978) test which will be studied extensively in Chapter 11. Davidson, Godfrey, and MacKinnon (1985) show that, like the Hausman test, the PSW test is equivalent to a much simpler omitted variables test, the omitted variables being the sum of the lagged and one-period ahead values of the regressors. Thus if the regression equation we are considering is

the PSW test involves estimating the expanded regression equation yt = β 1 x1t + β 2 x2t + γ 1 z1t + γ 2 z2t + ut

(8.56)

where z1t = x1,t+1 + x1,t−1 and z2t = x2,t+1 + x2,t−1 and testing the hypothesis γ 1 = γ 2 = 0 by the usual F -test. If there are lagged dependent variables in the equation, the test needs a minor modification. Suppose that the model is yt = β 1 yt−1 + β 2 xt + ut

(8.57)

Now the omitted variables would be defined as z1t = yt + yt−2 and z2t = xt+1 + xt−1 . There is no problem with z2t but z1t would be correlated with the error term ut because of the presence of yt in it. The solution would be simply to transfer it to the left hand side and write the expanded regression equation in (8.56) as (1 − γ 1 )yt = β 1 yt−1 + β 2 xt + γ 1 yt−2 + γ 2 z2t + ut

(8.58)


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197

This equation can be written as yt = β ∗1 yt−1 + β ∗2 xt + γ ∗1 yt−2 + γ ∗2 z2t + u∗t

(8.59)

where all the starred parameters are the corresponding unstarred ones divided by (1 − γ 1 ). The PSW now tests the hypothesis γ ∗1 = γ ∗2 = 0. Thus, in the case where the model involves the lagged dependent variable yt−1 as an explanatory variable, the only modification needed is that we should use yt−2 as the omitted variable, not (yt + yt−2 ). Note that it is only yt−1 that creates a problem, not higher-order lags of yt , like yt−2 , yt−3 , and so on. For yt−2 , the corresponding zt will be obtained by adding yt−1 to yt−3 . This zt is not correlated with ut as long as the disturbances are not serially correlated. (4) Tests for Non-nested Hypothesis Consider the following two competing non-nested models: H1 ; y = X1 β 1 + ǫ1

(8.60)

H2 ; y = X2 β 2 + ǫ2

(8.61)

These are non-nested because the explanatory variables under one model are not a subset of the other model even though X1 and X2 may share some common variables. In order to test H1 versus H2 , Cox (1961) modified the LR-test to allow for the non-nested case. The idea behind Cox’s approach is to consider to what extent Model I under H1 , is capable of predicting the performance of Model II, under H2 . Alternatively, one can artificially nest the 2 models H3 ; y = X1 β 1 + X2∗ β ∗2 + ǫ3

(8.62)

where X2∗ excludes from X2 the common variables with X1 . A test for H1 is simply the F -test for H0 ; β ∗2 = 0. Criticism: This tests H1 versus H3 which is a (Hybrid) of H1 and H2 and not H1 versus H2 . Davidson and MacKinnon (1981) proposed (testing α = 0) in the linear combination of H1 and H2 : y = (1 − α)X1 β 1 + αX2 β 2 + ǫ

(8.63)

′ −1  where α is an unknown scalar. Since α is not identified, we replace β 2 by β 2,OLS = (X2 X2 /T ) ′ (X2 y/T ) the regression coefficient estimate obtained from running y on X2 under H2 , i.e., (1) 2,OLS ; (2) Run y on X1 and y2 and test that the coefficient of y2 is Run y on X2 get yˆ2 = X2 β zero. This is known as the J-test and this is asymptotically N (0, 1) under H1 . Fisher and McAleer (1981) suggested a modification of the J-test known as the JA test.

 = plim(X ′ X2 /T )−1 plim(X ′ X1 /T )β + 0 Under H1 ; plimβ 2 1 2 2

(8.64)

′ −1  by β  = (X ′ X2 )−1 (X ′ X1 )β   Therefore, they propose replacing β 2 2 1,OLS where β 1,OLS = (X1 X1 ) 2 2 ′ X1 y. The steps for the JA-test are as follows:


198

Chapter 8: Regression Diagnostics and Specification Tests

 1,OLS . 1. Run y on X1 get y1 = X1 β

2. Run y1 on X2 get y2 = X2 (X2′ X2 )−1 X2′ y1 .

3. Run y on X1 and y2 and test that the coefficient of y2 is zero. This is the simple t-statistic on the coefficient of y2 . The J and JA tests are asymptotically equivalent.

Criticism: Note the asymmetry of H1 and H2 . Therefore one should reverse the role of these hypotheses and test again. In this case one can get the four scenarios depicted in Table 8.5. In case both hypotheses are not rejected, the data are not rich enough to discriminate between the two hypotheses. In case both hypotheses are rejected neither model is useful in explaining the variation in y. In case one hypothesis is rejected while the other is not, one should remember that the non-rejected hypothesis may still be brought down by another challenger hypothesis. Small Sample Properties: (i) The J-test tends to reject the null more frequently than it should. Also, the JA test has relatively low power when K1 , the number of parameters in H1 is larger than K2 , the number of parameters in H2 . Therefore, one should use the JA test when K1 is about the same size as K2 , i.e., the same number of non-overlapping variables. (ii) If both H1 and H2 are false, these tests are inferior to the standard diagnostic tests. In practice, use higher significance levels for the J-test, and supplement it with the artificially nested F -test and standard diagnostic tests. Table 8.5 Non-nested Hypothesis Testing α=0

α=1

Not Rejected

Rejected

Not Rejected

Both H1 and H2 are not rejected

H1 rejected H2 not rejected

Rejected

H1 not rejected H2 rejected

Both H1 and H2 are rejected

Note: J and JA tests are one degree of freedom tests, whereas the artificially nested F -test is not. For a recent summary of non-nested hypothesis testing, see Pesaran and Weeks (2001). Examples of non-nested hypothesis encountered in empirical economic research include linear versus log-linear models, see section 8.5. Also, logit versus probit models in discrete choice, see Chapter 13 and exponential versus Weibull distributions in the analysis of duration data. In the logit versus probit specification, the set of regressors is most likely to be the same. It is only the form of the distribution functions that separate the two models. Pesaran and Weeks (2001, p. 287) emphasize the differences between hypothesis testing and model selection: The model selection process treats all models under consideration symmetrically, while hypothesis testing attributes a different status to the null and to the alternative hypotheses and by design treats the models asymmetrically. Model selection always


8.3

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199

ends in a definite outcome, namely one of the models under consideration is selected for use in decision making. Hypothesis testing on the other hand asks whether there is any statistically significant evidence (in the Neyman-Pearson sense) of departure from the null hypothesis in the direction of one or more alternative hypotheses. Rejection of the null hypothesis does not necessarily imply acceptance of any one of the alternative hypotheses; it only warns the investigator of possible shortcomings of the null that is being advocated. Hypothesis testing does not seek a definite outcome and if carried out with due care need not lead to a favorite model. For example, in the case of nonnested hypothesis testing it is possible for all models under consideration to be rejected, or all models to be deemed as observationally equivalent. They conclude that the choice between hypothesis testing and model selection depends on the primary objective of one’s study. Model selection may be more appropriate when the objective is decision making, while hypothesis testing is better suited to inferential problems. A model may be empirically adequate for a particular purpose, but of little relevance for another use... In the real world where the truth is elusive and unknowable both approaches to model evaluation are worth pursuing.

(5) White’s (1982) Information-Matrix (IM) Test This is a general specification test much like the Hausman (1978) specification test which will be considered in details in Chapter 11. The latter is based on two different estimates of the regression coefficients, while the former is based on two different estimates of the Information Matrix I(θ) where θ ′ = (β ′ , σ 2 ) in the case of the linear regression studied in Chapter 7. The first estimate of I(θ) evaluates the expectation of the second derivatives of the log-likelihood at the MLE, i.e.,  −E(∂ 2 logL/∂θ∂θ′ ) at  θmle while the second sum up the outer products of the score vectors ni=1 (∂logLi (θ)/∂θ)(∂logLi (θ)/∂θ)′ evaluated at  θmle . This is based on the fundamental identity that I(θ) = −E(∂ 2 logL/∂θ∂θ′ ) = E(∂logL/∂θ)(∂logL/∂θ)′

If the model estimated by MLE is not correctly specified, this equality will not hold. From Chapter 7, equation (7.19), we know that for the linear regression model with normal disturbances, the first estimate of I(θ) denoted by I1 ( θ mle ) is given by

′ X X/ σ2 0 (8.65) I1 ( θM LE ) = 0 n/2 σ4 where σ 2 = e′ e/n is the MLE of σ 2 and e denotes the OLS residuals. Similarly, one can show that the second estimate of I(θ) denoted by I2 (θ) is given by    n (∂logLi (θ) ′ i=1 (∂logLi (θ) I2 (θ) = ∂θ ∂θ ⎤ ⎡ 2 ′ −ui xi u3i xi ui xi xi + n ⎢ σ4 3 ′ 2σ 4 2 2σ 6 4 ⎥ (8.66) = ′ i=1 ⎣ 1 u x ui ui ⎦ ui x − + − 4i + i 6i 2σ 2σ 4σ 4 2σ 6 4σ 8


200

Chapter 8: Regression Diagnostics and Specification Tests

where xi is the i-th row of X. Substituting the MLE we get n 3 ⎤ ⎡ n 2 ′ i=1 ei xi i=1 ei xi xi 6 ⎥ ⎢ 4 3 ′ 2 σ (8.67) θM LE ) = ⎣ n σ I2 ( n 4 ⎦ n i=1 ei i=1 ei xi − + 2 σ6 4 σ4 4 σ8  where we used the fact that ni=1 ei xi = 0. If the model is correctly specified and the disturbances are normal then plim I1 ( θM LE )/n = plim I2 ( θM LE )/n = I(θ)

Therefore, the Information Matrix (IM) test rejects the model when [I2 ( θM LE ) − I1 ( θM LE )]/n

(8.68)

is too large. These are two matrices with (k + 1) by (k + 1) elements since β is k × 1 and σ 2 is a scalar. However, due to symmetry, this reduces to (k + 2)(k + 1)/2 unique elements. Hall (1987) noted that the first k(k elements obtained from the first k × k block of + 1)/2 unique (8.68) have a typical element ni=1 (e2i − σ 2 )xir xis /n σ 4 where r and s denote the r-th and s-th explanatory variables with r, s = 1, 2, . . . , k. This term measures the discrepancy between the  OLS estimates of the variance-covariance matrix of β OLS and its robust counterpart suggested by White (1980), see Chapter 5. The next k unique elements correspond to the off-diagonal block n 3 6  σ e x /2n σ and this measures the discrepancy between the estimates of the cov(β, 2 ). The i=1 i i last element correspond to the difference in the bottom right elements, i.e., the two estimates of σ 2 . This is given by

1 n 4 3 8 e /4 σ − 4+ n i=1 i 4 σ These (k +1)(k +2)/2 unique elements can be arranged in vector form D(θ) which has a limiting normal distribution with zero mean and some covariance matrix V (θ) under the null. One can show, see Hall (1987) or Kr¨ amer and Sonnberger (1986) that if V (θ) is estimated from the sample moments of these terms, that the IM test statistic is given by H

m = nD ′ (θ)[V (θ)]−1 D(θ) →0 χ2(k+1)(k+2)/2

(8.69)

In fact, Hall (1987) shows that this statistic is the sum of three asymptotically independent terms m = m1 + m2 + m3

(8.70)

where m1 = a particular version of White’s heteroskedasticity test; m2 = n times the explained σ 6 ; and sum of squares from the regression of e3i on xi divided by 6 m3 =

2 n n 4 σ4 i=1 ei /n − 3 8 24 σ

which is similar to the Jarque-Bera test for normality of the disturbances given in Chapter 5. It is clear that the IM test will have power whenever the disturbances are non-normal or heteroskedastic. However, Davidson and MacKinnon (1992) demonstrated that the IM test


8.3

Specification Tests

201

considered above will tend to reject the model when true, much too often, in finite samples. This problem gets worse as the number of degrees of freedom gets large. In Monte Carlo experiments, Davidson and MacKinnon (1992) showed that for a linear regression model with ten regressors, the IM test rejected the null at the 5% level, 99.9% of the time for n = 200. This problem did not disappear when n increased. In fact, for n = 1000, the IM test still rejected the null 92.7% of the time at the 5% level. These results suggest that it may be more useful to run individual tests for non-normality, heteroskedasticity and other misspecification tests considered above rather than run the IM test. These tests may be more powerful and more informative than the IM test. Alternative methods of calculating the IM test with better finite-sample properties are suggested in Orme (1990), Chesher and Spady (1991) and Davidson and MacKinnon (1992). Example 3: For the consumption-income data given in Table 5.1, we first compute the RESET test from the consumption-income regression given in Chapter 5. Using EViews, one clicks on stability tests and then selects RESET. You will be prompted with the option of the number of fitted terms to include (i.e., powers of y). Table 8.6 shows the RESET test including y2 and y3 . The F -statistic for their joint-significance is equal to 40.98. This is significant and indicates misspecification. Next, we compute Utts (1982) Rainbow test. Table 8.7 gives the middle half of the sample, i.e., 1962-1983, and the SAS regression using this data. The RSS of these middle observations is given by e′ e = 178245.867, while the RSS for the entire sample is given by e′ e = 990923.131 so that the observed F -statistic given in (8.49) can be computed as follows: F =

(990923.131 − 178245.867)/22 = 4.145 178245.867/20

Table 8.6 Ramsey RESET Test F-statistic Log likelihood ratio Test Equation: Dependent Variable: Method: Sample: Included observations:

40.97877 49.05091

Probability Probability

0.00000 0.00000

CONSUM Least Squares 1950 1993 44

Variable C Y FITTEDˆ2 FITTEDˆ3

Coefficient –3210.837 2.079184 –0.000165 6.66E–09

Std. Error 1299.234 0.395475 4.71E–05 1.66E–09

t-Statistic –2.471330 5.257428 –3.494615 4.012651

R-squared Adjusted R-squared S.E. of regression Sum squared resid Log likelihood Durbin-Watson stat

0.998776 0.998684 90.13960 325005.9 –258.3963 1.458024

Mean dependent var S.D. dependent var Akaike info criterion Schwarz criterion F-statistic Prob (F-statistic)

Prob. 0.0178 0.0000 0.0012 0.0003 9250.54 2484.62 11.9271 12.0893 10876.9 0.00000


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Chapter 8: Regression Diagnostics and SpeciďŹ cation Tests

Table 8.7 Utts (1982) Rainbow Test OBS

YEAR

INCOME

CONSUMP

LEVERAGE

1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22

1962 1963 1964 1965 1966 1967 1968 1969 1970 1971 1972 1973 1974 1975 1976 1977 1978 1979 1980 1981 1982 1983

7583 7718 8140 8508 8822 9114 9399 9606 9875 10111 10414 11013 10832 10906 11192 11406 11851 12039 12005 12156 12146 12349

6931 7089 7384 7703 8005 8163 8506 8737 8842 9022 9425 9752 9602 9711 10121 10425 10744 10876 10746 10770 10782 11179

0.0440 0.0419 0.0359 0.0315 0.0285 0.0263 0.0246 0.0238 0.0230 0.0227 0.0229 0.0250 0.0241 0.0244 0.0260 0.0275 0.0316 0.0337 0.0333 0.0352 0.0350 0.0377

Dependent Variable: CONSUMPTION Analysis of Variance Source

DF

Model Error C Total

1 20 21

Root MSE Dep Mean C.V.

94.40494 9296.13636 1.01553

Sum of Squares 37715232.724 178245.86703 37893478.591

Mean Square 37715232.724 8912.29335

R-square Adj R-sq

F Value

Prob>F

4231.821

0.0001

0.9953 0.9951

Parameter Estimates

Variable INTERCEP INCOME

DF

Parameter Estimate

1 1

290.338091 0.872098

Standard Error 139.89449574 0.01340607

T for H0: Parameter=0 2.075 65.052

Prob > |T| 0.0511 0.0001


8.3

Specification Tests

203

Table 8.8 PSW Differencing Test Dependent Variable: Method: Sample (adjusted): Included observations:

CONSUM Least Squares 1951 1993 43 after adjusting endpoints

Variable C Y Z

Coefficient –60.28744 0.501803 0.208394

Std. Error 94.40916 0.273777 0.137219

t-Statistic –0.638576 1.832891 1.518701

R-squared Adjusted R-squared S.E. of regression Sum squared resid Log likelihood Durbin-Watson stat

0.996373 0.996191 151.5889 919167.2 –275.3699 0.456301

Mean dependent var S.D. dependent var Akaike info criterion Schwarz criterion F-statistic Prob (F-statistic)

Prob. 0.5267 0.0743 0.1367 9330.326 2456.343 12.94744 13.07031 5493.949 0.00000

This is distributed as F22,20 under the null hypothesis and rejects the hypothesis of linearity.   The PSW differencing test, given in (8.54), is based upon β OLS = 0.915623 with V (β OLS ) =   0.0000747926, and β  = (0.843298 − F D = 0.843298 with V (β F D ) = 0.00732539. Therefore, q 0.915623) = −0.072325 with V ( q )/T = 0.00732539 − 0.0000747926 = 0.0072506 and Δ = T q′ [V ( q )]−1 q = 0.721 which is distributed as χ21 under the null. This is not significant and does not reject the null of no misspecification. The artificial regression given in (8.56) with Zt = Yt + Yt−1 is given in Table 8.8. The t-statistic for Zt is 1.52 and has a p-value of 0.137 which is not significant. Now consider the two competing non-nested models: H1 ; Ct = β 0 + β 1 Yt + β 2 Yt−1 + ut

H2 ; Ct = γ 0 + γ 1 Yt + γ 2 Ct−1 + vt

The two non-nested models share Yt as a common variable. The artificial model that nests these two models is given by: H3 ; Ct = δ0 + δ 1 Yt + δ 2 Yt−1 + δ 3 Ct−1 + ǫt 2 (C 2HAT ). RegresTable 8.9, runs regression (1) given by H2 and obtains the predicted values C sion (2) runs consumption on a constant, income, lagged income and C 2HAT. The coefficient of this last variable is 1.64 and is statistically significant with a t-value of 7.80. This is the Davidson and MacKinnon (1981) J-test. In this case, H1 is rejected but H2 is not rejected. The JA-test, given by Fisher and McAleer (1981) runs the regression in H1 and keeps the predicted 1 (C 1HAT ). This is done in regression (3). Then C 1HAT is run on a constant, income values C 2 (C 2TILDE ). This is done and lagged consumption and the predicted values are stored as C in regression (5). The last step runs consumption on a constant, income, lagged income and C 2TILDE, see regression (6). The coefficient of this last variable is 6.53 and is statistically significant with a t-value of 7.80. Again H1 is rejected but H2 is not rejected. Reversing the roles of H1 and H2 , the J and JA-tests are repeated. In fact, regression (4) 1 (which was obtained from runs consumption on a constant, income, lagged consumption and C


204

Chapter 8: Regression Diagnostics and Specification Tests

1 is −2.54 and is statistically significant with a t-value of regression (3)). The coefficient on C 2 on a constant, −4.13. This J-test rejects H2 but does not reject H1 . Regression (7) runs C  income and lagged income and the predicted values are stored as C1 (C1TILDE). The last step of 1 , see regression the JA test runs consumption on a constant, income, lagged consumption and C (8). The coefficient of this last variable is −1.18 and is statistically significant with a t-value of −4.13. This JA test rejects H2 but not H1 . The artificial model, given in H3 , is also estimated, see regression (9). One can easily check that the corresponding F -tests reject H1 against H3 and also H2 against H3 . In sum, all evidence indicates that both Ct−1 and Yt−1 are important to include along with Yt . Of course, the true model is not known and could include higher lags of both Yt and Ct .

8.4

Nonlinear Least Squares and the Gauss-Newton Regression4

So far we have been dealing with linear regressions. But, in reality, one might face a nonlinear regression of the form: yt = xt (β) + ut

for t = 1, 2, . . . , T

(8.71)

where ut ∼ IID(0, σ 2 ) and xt (β) is a scalar nonlinear regression function of k unknown parameters β. It can be interpreted as the expected value ofyt conditional on the values of the independent variables. Nonlinear least squares minimizes Tt=1 (yt − xt (β))2 = (y − x(β))′ (y − x(β)). The first-order conditions for minimization yield   =0 X ′ (β)(y − x(β))

(8.72)

where X(β) is a T × k matrix with typical element Xtj (β) = ∂xt (β)/∂β j for j = 1, . . . , k. The solution to these k equations yield the Nonlinear Least Squares (NLS) estimates of β denoted by  β N LS . These normal equations given in (8.72) are similar to those in the linear case in that they  to be orthogonal to the matrix of derivatives X(β).  In the require the vector of residuals y −x(β)     linear case, x(β) = X β OLS and X(β) = X where the latter is independent of β. Because of this  as well as the matrix of derivatives X(β)  on β,  one in general dependence of the fitted values x(β) cannot get explicit analytical solution to these NLS first-order equations. Under fairly general  conditions, see Davidson and MacKinnon (1993), one can show that the β N LS has asymptotically 2 ′ a normal distribution with mean β 0 and asymptotic variance σ 0 (X (β 0 )X(β 0 ))−1 , where β 0 and σ 0 are the true values of the parameters generating the data. Similarly, defining ′   s2 = (y − x(β N LS )) (y − x(β N LS ))/(T − k)

  −1 . If the disturbances we get a feasible estimate of this covariance matrix as s2 (X ′ (β)X( β)) are normally distributed then NLS is MLE and therefore asymptotically efficient as long as the model is correctly specified, see Chapter 7. Taking the first-order Taylor series approximation around some arbitrary parameter vector β ∗ , we get y = x(β ∗ ) + X(β ∗ )(β − β ∗ ) + higher-order terms + u

(8.73)

y − x(β ∗ ) = X(β ∗ )b + residuals

(8.74)

or


8.4

Nonlinear Least Squares and the Gauss-Newton Regression

205

Table 8.9 Non-nested J and JA Test Regression 1 Dependent Variable: C Analysis of Variance Source Model Error C Total Root MSE Dep Mean C.V.

DF 2 40 42 113.51027 9330.32558 1.21657

Sum of Squares 252896668.15 515383.28727 253412051.44 R-square Adj R-sq

Mean Square 126448334.08 12884.58218

F Value 9813.926

Prob>F 0.0001

Standard Error 70.43957345 0.07458032 0.08203649

T for H0: Parameter=0 –0.248 6.369 5.954

Prob > |T| 0.8051 0.0001 0.0001

Mean Square 84351096.151 9199.05101

F Value 9169.543

Prob>F 0.0001

T for H0: Parameter=0 –0.202 –0.446 –4.126 7.805

Prob > |T| 0.8409 0.6579 0.0002 0.0001

0.9980 0.9979

Parameter Estimates Variable INTERCEP Y CLAG

DF 1 1 1

Parameter Estimate –17.498706 0.475030 0.488458

DF 3 39 42

Sum of Squares 253053288.45 358762.98953 253412051.44

Regression 2 Dependent Variable: C Analysis of Variance Source Model Error C Total Root MSE Dep Mean C.V.

95.91168 9330.32558 1.02796

R-square Adj R-sq

0.9986 0.9985

Parameter Estimates

Variable INTERCEP Y YLAG C2HAT

DF 1 1 1 1

Parameter Estimate –12.132250 –0.058489 –0.529709 1.637806

Standard Error 60.05133697 0.13107211 0.12837635 0.20983763


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Chapter 8: Regression Diagnostics and Specification Tests

Table 8.9 (continued) Regression 3 Dependent Variable: C Analysis of Variance Source Model Error C Total Root MSE Dep Mean C.V.

DF 2 40 42 151.58885 9330.32558 1.62469

Sum of Squares 252492884.22 919167.21762 253412051.44 R-square Adj R-sq

Mean Square 126246442.11 22979.18044

F Value 5493.949

Prob>F 0.0001

Standard Error 94.40916038 0.13669864 0.13721871

T for H0: Parameter=0 –0.639 5.195 1.519

Prob > |T| 0.5267 0.0001 0.1367

Mean Square 84351096.151 9199.05101

F Value 9169.543

Prob>F 0.0001

T for H0: Parameter=0 –2.647 5.042 7.805 –4.126

Prob > |T| 0.0117 0.0001 0.0001 0.0002

0.9964 0.9962

Parameter Estimates Variable INTERCEP Y YLAG

DF 1 1 1

Parameter Estimate –60.287436 0.710197 0.208394

DF 3 39 42

Sum of Squares 253053288.45 358762.98953 253412051.44

Regression 4 Dependent Variable: C Analysis of Variance Source Model Error C Total Root MSE Dep Mean C.V.

95.91168 9330.32558 1.02796

R-square Adj R-sq

0.9986 0.9985

Parameter Estimates

Variable INTERCEP Y CLAG C1HAT

DF 1 1 1 1

Parameter Estimate –194.034085 2.524742 0.800000 –2.541862

Standard Error 73.30021926 0.50073386 0.10249693 0.61602660


8.4

Nonlinear Least Squares and the Gauss-Newton Regression

207

Table 8.9 (continued) Regression 5 Dependent Variable: C1HAT Analysis of Variance Source Model Error C Total Root MSE Dep Mean C.V.

DF 2 40 42 24.61739 9330.32558 0.26384

Sum of Squares 252468643.58 24240.64119 252492884.22 R-square Adj R-sq

Mean Square 126234321.79 606.01603

F Value 208301.952

Prob>F 0.0001

Standard Error 15.27649082 0.01617451 0.01779156

T for H0: Parameter=0 –4.546 49.855 6.889

Prob > |T| 0.0001 0.0001 0.0001

Mean Square 84351096.151 9199.05101

F Value 9169.543

Prob>F 0.0001

T for H0: Parameter=0 4.849 –6.695 –4.126 7.805

Prob > |T| 0.0001 0.0001 0.0002 0.0001

0.9999 0.9999

Parameter Estimates

Variable INTERCEP Y CLAG

DF 1 1 1

Parameter Estimate –69.451205 0.806382 0.122564

DF 3 39 42

Sum of Squares 253053288.45 358762.98953 253412051.44

Regression 6 Dependent Variable: C Analysis of Variance Source Model Error C Total Root MSE Dep Mean C.V.

95.91168 9330.32558 1.02796

R-square Adj R-sq

0.9986 0.9985

Parameter Estimates

Variable INTERCEP Y YLAG C2TILDE

DF 1 1 1 1

Parameter Estimate 412.528849 –4.543882 –0.529709 6.527181

Standard Error 85.07504557 0.67869220 0.12837635 0.83626992


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Chapter 8: Regression Diagnostics and Specification Tests

Table 8.9 (continued) Regression 7 Dependent Variable: C2HAT Analysis of Variance Source Model Error C Total Root MSE Dep Mean C.V.

DF 2 40 42 72.27001 9330.32558 0.77457

Sum of Squares 252687749.95 208918.20026 252896668.15 R-square Adj R-sq

Mean Square 126343874.98 5222.95501

F Value 24190.114

Prob>F 0.0001

Standard Error 45.00958513 0.06517110 0.06541905

T for H0: Parameter=0 –0.653 7.202 6.889

Prob > |T| 0.5173 0.0001 0.0001

Mean Square 84351096.151 9199.05101

F Value 9169.543

Prob>F 0.0001

T for H0: Parameter=0 –1.232 6.263 7.805 –4.126

Prob > |T| 0.2252 0.0001 0.0001 0.0002

0.9992 0.9991

Parameter Estimates

Variable INTERCEP Y YLAG

DF 1 1 1

Parameter Estimate –29.402245 0.469339 0.450666

DF 3 39 42

Sum of Squares 253053288.45 358762.98953 253412051.44

Regression 8 Dependent Variable: C Analysis of Variance Source Model Error C Total Root MSE Dep Mean C.V.

95.91168 9330.32558 1.02796

R-square Adj R-sq

0.9986 0.9985

Parameter Estimates

Variable INTERCEP Y CLAG C1TILDE

DF 1 1 1 1

Parameter Estimate –75.350919 1.271176 0.800000 –1.175392

Standard Error 61.14775168 0.20297803 0.10249693 0.28485929


8.4

Nonlinear Least Squares and the Gauss-Newton Regression

209

Table 8.9 (continued) Regression 9 Dependent Variable: C Analysis of Variance Source Model Error C Total Root MSE Dep Mean C.V.

DF 3 39 42 95.91168 9330.32558 1.02796

Sum of Squares 253053288.45 358762.98953 253412051.44 R-square Adj R-sq

Mean Square 84351096.151 9199.05101

F Value 9169.543

Prob>F 0.0001

T for H0: Parameter=0 –0.682 8.318 –4.126 7.805

Prob > |T| 0.4991 0.0001 0.0002 0.0001

0.9986 0.9985

Parameter Estimates Variable INTERCEP Y YLAG CLAG

DF 1 1 1 1

Parameter Estimate –40.791743 0.719519 –0.529709 0.800000

Standard Error 59.78575886 0.08649875 0.12837635 0.10249693

This is the simplest version of the Gauss-Newton Regression, see Davidson and MacKinnon (1993). In this case the higher-order terms and the error term are combined in the residuals and (β − β ∗ ) is replaced by b, a parameter vector that can be estimated. If the model is linear, N LS , the X(β ∗ ) is the matrix of regressors X and the GNR regresses a residual on X. If β ∗ =β unrestricted NLS estimator of β, then the GNR becomes  + residuals y−x  = Xb

(8.75)

   where x  ≡ x(β N LS ) and X ≡ X(β N LS ). From the first-order conditions of NLS we get (y −  = 0. In this case, OLS on this GNR yields bOLS = (X  ′ X)  −1 X  ′ (y − x x )′ X ) = 0 and this GNR has no explanatory power. However, this regression can be used to (i) check that the firstorder conditions given in (8.72) are satisfied. For example, one could check that the t-statistics are of the 10−3 order, and that R2 is zero up to several decimal places; (ii) compute estimated  ′ X)  −1 , where s2 = (y − x covariance matrices. In fact, this GNR prints out s2 (X )′ (y − x )/(T − k) is the OLS estimate of the regression variance. This can be verified easily using the fact that this GNR has no explanatory power. This method of computing the estimated variance-covariance  has been obtained by some method other than NLS. matrix is useful especially in cases where β For example, sometimes the model is nonlinear only in one or two parameters which are known to be in a finite range, say between zero and one. One can then search over this range, running OLS regressions and minimizing the residual sum of squares. This search procedure can be repeated over finer grids to get more accuracy. Once the final parameter estimate is found, one can run the GNR to get estimates of the variance-covariance matrix.


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Chapter 8: Regression Diagnostics and Specification Tests

Testing Restrictions (GNR Based on the Restricted NLS Estimates) The best known use for the GNR is to test restrictions. These are based on the LM principle which requires only the restricted estimator. In particular, consider the following competing hypotheses: H0 ; y = x(β 1 , 0) + u

H1 ; y = x(β 1 , β 2 ) + u

 the restricted where u ∼ IID(0, σ 2 I) and β 1 and β 2 are k ×1 and r ×1, respectively. Denote by β ′ ′  , 0).  = (β NLS estimator of β, in this case β 1 The GNR evaluated at this restricted NLS estimator of β is  1 b1 + X 2 b2 + residuals (y − x ) = X

(8.76)

2 b2 + residuals P¯X1 (y − x ) = P¯X 1 X

(8.77)

2 b2 + residuals (y − x ) = P¯X 1 X

(8.78)

 −1  ′  ′ P¯  X  −1  ′ ¯  (y − x  ′ P¯  X ) ) = (X b2,OLS = (X 2 X1 2 ) X2 (y − x 2 X1 2 ) X2 PX 1

(8.79)

 with Xi (β) = ∂x/∂β i for i = 1, 2.  and X i = Xi (β) where x  = x(β) By the FWL Theorem this yields the same estimate of b2 as

 ′ (y − x ) = (y − x ) since X ) = 0 from the first-order ) = (y − x ) − PX 1 (y − x But P¯X 1 (y − x 1 conditions of restricted NLS. Hence, (8.77) reduces to Therefore,

2 (X  ′ P¯  X  −1  ′ and the residual sums of squares is (y − x )′ (y − x ) − (y − x )′ X ). 2 X1 2 ) X2 (y − x ′  If X2 was excluded from the regression in (8.76), (y − x ) (y − x ) would be the residual sum of squares. Therefore, the reduction in the residual sum of squares brought about by the inclusion 2 is of X  −1  ′ 2 (X  ′ P¯  X ) (y − x )′ X 2 X1 2 ) X2 (y − x

1 has no explanatory This is also equal to the explained sum of squares from (8.76) since X 2 power. This sum of squares divided by a consistent estimate of σ is asymptotically distributed as χ2r under the null. Different consistent estimates of σ 2 yield different test statistics. The two most common test statistics for H0 based on this regression are the following: (1) T Ru2 where Ru2 is the uncentered R2 of (8.76) and (2) the F -statistic for b2 = 0. The first statistic is given by T Ru2 =  −1  ′ 2 (X  ′ P¯  X )/(y − x )′ (y − x ) where the uncentered R2 was defined in the T (y − x )′ X 2 X1 2 ) X2 (y − x Appendix to Chapter 3. This statistic implicitly divides the explained sum of squares term by σ 2 = (restricted residual sums of squares)/T . This is equivalent to the LM-statistic obtained  and getting the explained sum of squares. by running the artificial regression (y − x )/ σ on X 2 Regression packages print the centered R . This is equal to the uncentered Ru2 as long as there is a constant in the restricted regression so that (y − x ) sum to zero.


8.4

Nonlinear Least Squares and the Gauss-Newton Regression

211

The F -statistic for b2 = 0 from (8.76) is 2 (X  ′ P¯  X  −1  ′ (y − x )′ X )/r (RRSS − URSS )/r 2 X1 2 ) X2 (y − x = (8.80) ′ ′ ′ ′ −1 2 (X  P¯  X   URSS /(T − k) [(y − x ) (y − x ) − (y − x ) X )]/(T − k) 2 X1 2 ) X2 (y − x

The denominator is the OLS estimate of σ 2 from (8.76) which tends to σ 20 as T → ∞. Hence (rF -statistic → χ2r ). In small samples, use the F -statistic. Diagnostic Tests for Linear Regression Models Variable addition tests suggested by Pagan and Hall (1983) consider the additional variables Z of dimension (T × r) and test whether their coefficients are zero using an F -test from the regression y = Xβ + Zγ + u

(8.81)

If H0 ; γ = 0 is true, the model is y = Xβ + u and there is no misspecification. The GNR for this restriction would run the following regression: P¯X y = Xb + Zc + residuals

(8.82)

and test that c is zero. By the FWL Theorem, (8.82) yields the same residual sum of squares as P¯X y = P¯X Zc + residuals

(8.83)

Applying the FWL Theorem to (8.81) we get the same residual sum of squares as the regression in (8.83). The F -statistic for γ = 0 from (8.81) is therefore identical to the F -statistic for c = 0 from the GNR given in (8.82). Hence, “Tests based on the GNR are equivalent to variable addition tests when the latter are applicable,” see Davidson and MacKinnon (1993, p. 194). Note also, that the nRu2 test statistic for H0 ; γ = 0 based on the GNR in (8.82) is exactly the LM statistic based on running the restricted least squares residuals of y on X on the unrestricted set of regressors X and Z in (8.81). If X has a constant, then the uncentered R2 is equal to the centered R2 printed by the regression. Computational Warning: It is tempting to base tests on the OLS residuals u  = P¯X y by simply regressing them on the test regressors Z. This is equivalent to running the GNR without the X variables on the right hand side of (8.82) yielding test-statistics that are too small. Functional Form Davidson and MacKinnon (1993, p. 195) show that the RESET with yt = Xt β + yt2 c+ residual which is based on testing for c = 0 is equivalent to testing for θ = 0 using the nonlinear model yt = Xt β(1 + θXt β) + ut . In this case, it is easy to verify from (8.74) that the GNR is yt − Xt β(1 + θXt β) = (2θ(Xt β)Xt + Xt )b + (Xt β)2 c + residual

2    At θ = 0 and β = β OLS , the GNR becomes (yt − Xt β OLS ) = Xt b + (Xt β OLS ) c+ residual. The t-statistic on c = 0 is equivalent to that from the RESET regression given in section 8.3, see problem 25.


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Chapter 8: Regression Diagnostics and Specification Tests

Testing for Serial Correlation Suppose that the null hypothesis is the nonlinear regression model given in (8.71), and the alternative is the model yt = xt (β) + ν t with ν t = ρν t−1 + ut where ut ∼ IID(0, σ 2 ). Conditional on the first observation, the alternative model can be written as yt = xt (β) + ρ(yt−1 − xt−1 (β)) + ut The GNR test for H0 ; ρ = 0, computes the derivatives of this regression function with respect to β and ρ evaluated at the restricted estimates under the null hypothesis, i.e., ρ = 0 and  β=β N LS (the nonlinear least squares estimate of β assuming no serial correlation). Those yield  N LS ) and (yt−1 − xt−1 (β  N LS )) respectively. Therefore, the GNR runs u  N LS ) = t = yt − xt (β Xt (β  Xt (β N LS )b+ c ut−1 + residual, and tests that c = 0. If the regression model is linear, this reduces to running ordinary least squares residuals on their lagged values in addition to the regressors in the model. This is exactly the Breusch and Godfrey test for first-order serial correlation considered in Chapter 5. For other applications as well as benefits and limitations of the GNR, see Davidson and MacKinnon (1993).

8.5

Testing Linear versus Log-Linear Functional Form5

In many economic applications where the explanatory variables take only positive values, econometricians must decide whether a linear or log-linear regression model is appropriate. In general, the linear model is given by yi =

k

j=1 β j Xij

+

ℓ

s=1 γ s Zis

+ ui

i = 1, 2, . . . , n

(8.84)

+ ui

(8.85)

and the log-linear model is logyi =

k

j=1 β j

logXij +

ℓ

s=1 γ s Zis

i = 1, 2, . . . , n

with ui ∼ NID(0, σ 2 ). Note that, the log-linear model is general in that only the dependent variable y and a subset of the regressors, i.e., the X variables are subject to the logarithmic transformation. Of course, one could estimate both models and compare their log-likelihood values. This would tell us which model fits best, but not whether either is a valid specification. Box and Cox (1964) suggested the following transformation ⎧ λ ⎪ ⎨ yi λ−1 when λ = 0 (8.86) B(yi , λ) = ⎪ ⎩ logy when λ = 0 i where yi > 0. Note that for λ = 1, as long as there is constant in the regression, subjecting the linear model to a Box-Cox transformation is equivalent to not transformation yields the loglinear regression. Therefore, the following Box-Cox model regression. Therefore, the following Box-Cox model B(yi , λ) =

k

j=1 β j B(Xij , λ) +

ℓ

s=1 γ s Zis

+ ui

(8.87)


8.5

Testing Linear versus Log-Linear Functional Form

213

encompasses as special cases the linear and log-linear models given in (8.84) and (8.85), respectively. Box and Cox (1964) suggested estimating these models by ML and using the LR test to test (8.84) and (8.85) against (8.87). However, estimation of (8.87) is computationally burdensome, see Davidson and MacKinnon (1993). Instead, we give an LM test involving a Double Length Regression (DLR) due to Davidson and MacKinnon (1985) that is easier to compute. In fact, Davidson and MacKinnon (1993, p. 510) point out that “everything that one can do with the Gauss-Newton Regression for nonlinear regression models can be done with the DLR for models involving transformations of the dependent variable.” The GNR is not applicable in cases where the dependent variable is subjected to a nonlinear transformation, so one should use a DLR in these cases. Conversely, in cases where the GNR is valid, there is no need to run the DLR, since in these cases the latter is equivalent to the GNR. For the linear model (8.84), the null hypothesis is that λ = 1. In this case, Davidson and MacKinnon suggest running a regression with 2n observations where the dependent variable σ , . . . , en / σ , 1, . . . , 1)′ , i.e., the first n observations are the OLS residuals has observations (e1 / from (8.84) divided by the MLE of σ, where σ 2mle = e′ e/n. The second n observations are all equal to 1. The 2n observations for the regressors have typical elements: for for for for

β j : Xij − 1 for i = 1, . . . , n and 0 γ s : Zis for i = 1, . . . , n and 0 σ: ei / σ for i = 1, . . . , n and −1 k  λ: j=1 β j (Xij logXij − Xij + 1) − (yi log yi − yi+1 ) and σ logyi

for for for for for

the second n the second n the second n i = 1, . . . , n the second n

elements elements elements

β j : logXij for i = 1, . . . , n for i = 1, . . . , n γ s : Zis σ: ei / σ for i = 1, . . . , n 1 k  λ: 2 j=1 β j (logXij )2 − 12 (logyi )2

the second n elements the second n elements the second n elements

elements

The explained sum of squares for this DLR provides an asymptotically valid test for λ = 1. This will be distributed as χ21 under the null hypothesis. Similarly, when testing the log-linear model (8.85), the null hypothesis is that λ = 0. In this case, the dependent variable of the DLR has observations ( e1 / σ , e2 / σ , . . . , en / σ , 1, . . . , 1)′ , i.e., the first n observations are the OLS residuals from (8.85) divided by the MLE for σ, i.e., σ  where σ 2 = e′  e/n. The second n observations are all equal to 1. The 2n observations for the regressors have typical elements: for for for for

and 0 for and 0 for and −1 for for i = 1, . . . , n and σ logyi for

the second n elements

The explained sum of squares from this DLR provides an asymptotically valid test for λ = 0. This will be distributed as χ21 under the null hypothesis. For the cigarette data given in Table 3.2, the linear model is given by C = β 0 + β 1 P + β 2 Y + u whereas the log-linear model is given by logC = γ 0 + γ 1 logP + γ 2 logY + ǫ and the Box-Cox model is given by B(C, λ) = δ0 + δ 1 B(P, λ) + δ 2 B(Y, λ) + ν, where B(C, λ) is defined in (8.86). In this case, the DLR which tests the hypothesis that H0 ; λ = 1, i.e., the model is linear, gives an explained sum of squares equal to 15.55. This is greater than a χ21,0.05 = 3.84 and is therefore significant at the 5% level. Similarly the DLR that tests the hypothesis that H0 ; λ = 0, i.e., the model is log-linear, gives an explained sum of squares equal to 8.86. This is also greater than χ21,0.05 = 3.84 and is therefore significant at the 5% level. In this case, both the linear and log-linear models are rejected by the data.


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Chapter 8: Regression Diagnostics and Specification Tests

Finally, it is important to note that there are numerous other tests for testing linear and log-linear models and the interested reader should refer to Davidson and MacKinnon (1993).

Notes 1. This section is based on Belsley, Kuh and Welsch (1980). 2. Other residuals that are linear unbiased with a scalar covariance matrix (LUS) are the BLUS residuals suggested by Theil (1971). Since we are explicitly dealing with time-series data, we use subscript t rather than i to index observations and T rather than n to denote the sample size.

3. Ramsey’s (1969) initial formulation was based on BLUS residuals, but Ramsey and Schmidt (1976) showed that this is equivalent to using OLS residuals.

4. This section is based on Davidson and MacKinnon (1993, 2001). 5. This section is based on Davidson and MacKinnon (1993, pp. 502-510).

Problems 1. We know that H = PX is idempotent. Also, (In − PX ) is idempotent. Therefore, b′ Hb ≥ 0 for any arbitrary vector b. Using these facts, show for b′ = (1, 0, . . . , 0) that 0 ≤ h11 ≤ 1. Deduce that 0 ≤ hii ≤ 1 for i = 1, . . . , n. 2. For the simple regression with no constant yi = xi β + ui for i = 1, . . . , n n (a) What is hii ? Verify that i=1 hii = 1.

−β  , see (8.13)? What is s2 in terms of s2 and e2 , see (8.18)? What is DFBE(b) What is β (i) i (i) TAS ij , see (8.19)? (c) What are DFFIT i and DFFITS i , see (8.21) and (8.22)?

(d) What is Cook’s distance measure Di2 (s) for this simple regression with no intercept, see (8.24)? (e) Verify that (8.27) holds for this simple regression with no intercept. What is COVRATIO i , see (8.26)? 3. From the definition of s2(i) in (8.17), substitute (8.13) in (8.17) and verify (8.18). 4. Consider the augmented regression given in (8.5) y = Xβ ∗ + di ϕ + u where ϕ is a scalar and di = 1 for the i-th observation and 0 otherwise. Using the Frisch-Waugh Lovell Theorem given in section 7.3, verify that  ∗ = (X ′ X(i) )−1 X ′ y(i) = β  . (a) β (i) (i) (i)

(b) ϕ  = (d′i P¯X di )−1 d′i P¯X y = ei /(1 − hii ) where P¯X = I − PX .

(c) Residual Sum of Squares from (8.5) = (Residual Sum of Squares with di deleted) − e2i /(1 − hii ).

(d) Assuming Normality of u, show that the t-statistic for testing ϕ = 0 is t = ϕ  /s.e.( ϕ) = e∗i as given in (8.3).


Problems

215

5. Consider the augmented regression y = Xβ ∗ + P¯X Dp ϕ∗ + u, where Dp is an n × p matrix of dummy variables for the p suspected observations. Note that P¯X Dp rather than Dp appear in this equation. Compare with (8.6). Let ep = Dp′ e, then E(ep ) = 0, var(ep ) = σ 2 Dp′ P¯X Dp . Verify that  ∗ = (X ′ X)−1 X ′ y = β  (a) β OLS and ∗ ′ ¯ −1 ′ ¯ (b) ϕ  = (D PX Dp ) D PX y = (D′ P¯X Dp )−1 D′ e = (D′ P¯X Dp )−1 ep . p

p

p

p

p

(c) Residual Sum of Squares = (Residual Sum of Squares with Dp deleted) − e′p (Dp′ P¯X )Dp−1 ep . Using the Frisch-Waugh Lovell Theorem show this residual sum of squares is the same as that for (8.6).

(d) Assuming normality of u, verify (8.7) and (8.9). (e) Repeat this exercise for problem 4 with P¯X di replacing di . What do you conclude? 6. Using the updating formula in (8.11), verify (8.12) and deduce (8.13). 7. Verify that Cook’s distance measure given in (8.25) is related to DFFITS i (σ) as follows: DF√ FITS i (σ) = kDi (σ). 8. Using the matrix identity det(Ik − ab′ ) = 1 − b′ a, where a and b are column vectors of dimension ′ k, prove (8.27). Hint: Use a = xi and b′ = x′i (X ′ X)−1 and the fact that det[X(i) X(i) ] =det[{Ik − ′ ′ −1 ′ xi xi (X X) }X X]. 9. For the cigarette data given in Table 3.2 (a) Replicate the results in Table 8.2.  β  (b) For the New Hampshire observation (NH), compute  eN H , e∗N H , β− (N H) , DFBETAS N H , 2 DFFIT N H , DFFITS N H , DN H (s), COVRATIO N H , and FVARATIO N H . (c) Repeat the calculations in part (b) for the following states: AR, CT, NJ and UT.

(d) What about the observations for NV, ME, NM and ND? Are they influential? 10. For the Consumption-Income data given in Table 5.1, compute (a) The internal studentized residuals e given in (8.1).

(b) The externally studentized residuals e∗ given in (8.3). (c) Cook’s statistic given in (8.25).

(d) The leverage of each observation h. (e) The DFFITS given in (8.22). (f) The COVRATIO given in (8.28). (g) Based on the results in parts (a) to (f), identify the observations that are influential. 11. Repeat problem 10 for the 1982 data on earnings used in Chapter 4. This data is provided on the Springer web site as EARN.ASC. 12. Repeat problem 10 for the Gasoline data provided on the Springer web site as GASOLINE.DAT. Use the gasoline demand model given in Chapter 10, section 5. Do this for Austria and Belgium separately. 13. Independence of Recursive Residuals. (a) Using the updating formula given in (8.11) with A = (Xt′ Xt ) and a = −b = x′t+1 , verify (8.31).


216

Chapter 8: Regression Diagnostics and Specification Tests (b) Using (8.31), verify (8.32).

 (c) For ut ∼ IIN(0, σ 2 ) and wt+1 defined in (8.30) verify (8.33). Hint: define vt+1 = ft+1 wt+1 . From (8.30), we have   = x′ (β − β  ) + ut+1 for t = k, . . . , T − 1 vt+1 = ft+1 wt+1 = yt+1 − x′t+1 β t t+1 t Since ft+1 is fixed, it suffices to show that cov(vt+1 , vs+1 ) = 0 for t = s.

14. Recursive Residuals are Linear Unbiased With Scalar Covariance Matrix (LUS). (a) Verify that the (T − k) recursive residuals defined in (8.30) can be written in vector form as w = Cy where C is defined in (8.34). This shows that the recursive residuals are linear in y. (b) Show that C satisfies the three properties given in (8.35) i.e., CX = 0, CC ′ = IT −k , and C ′ C = P¯X . Prove that CX = 0 means that the recursive residuals are unbiased with zero mean. Prove that the CC ′ = IT −k means that the recursive residuals have a scalar covariance matrix. Prove that C ′ C = P¯X means that the sum of squares of (T − k) recursive residuals is equal to the sum of squares of T least squares residuals. (c) If the true disturbances u ∼ N (0, σ 2 IT ), prove that the recursive residuals w ∼ N (0, σ 2 IT −k ) using parts (a) and (b). 2 (d) Verify (8.36), i.e., show that RSSt+1 = RSSt + wt+1 for t = k, . . . , T − 1 where RSSt = ′   (Yt − Xt β t ) (Yt − Xt β t ).

15. The Harvey and Collier (1977) Misspecification t-test as a Variable Additions Test. This is based on Wu (1993). (a) Show that the F -statistic for testing H0 ; γ = 0 versus γ = 0 in (8.44) is given by F =

y ′ P¯X y − y ′ P¯[X,z] y ′ y P¯[X,z] y/(T − k − 1)

=

y ′ (P¯X

y ′ Pz y − Pz )y/(T − k − 1)

and is distributed as F (1, T − k − 1) under the null hypothesis.

(b) Using the properties of C given in (8.35), show that the F -statistic given in part (a) is the square of the Harvey and Collier (1977) t-statistic given in (8.43). 16. For the Gasoline data for Austria given on the Springer web site as GASOLINE.DAT and the model given in Chapter 10, section 5, compute: (a) The recursive residuals given in (8.30). (b) The CUSUM given in (8.46) and plot it against r. (c) Draw the 5% upper and lower lines given below (8.46) and see whether the CUSUM crosses these boundaries. (d) The post-sample predictive test for 1978. Verify that computing it from (8.38) or (8.40) yields the same answer. (e) The modified von Neuman ratio given in (8.42). (f) The Harvey and Collier (1977) functional misspecification test given in (8.43). 17. The Differencing Test in a Regression with Equicorrelated Disturbances. This is based on Baltagi (1990). Consider the time-series regression Y = ιT α + Xβ + u

(1)


Problems

217

where ιT is a vector of ones of dimension T . X is T ×K and [ιT , X] is of full column rank. u ∼ (0, Ω) where Ω is positive definite. Differencing this model, we get DY = DXβ + Du

(2)

where D is a (T − 1) × T matrix given below (8.50). Maeshiro and Wichers (1989) show that GLS on (1) yields through partitioned inverse:  = (X ′ LX)−1 X ′ LY β

(3)

 = (X ′ M X)−1 X ′ M Y β

(4)

where L = Ω−1 − Ω−1 ιT (ι′T Ω−1 ιT )−1 ι′T Ω−1 . Also, GLS on (2) yields

where M = D′ (DΩD′ )−1 D. Finally, they show that M = L, and GLS on (2) is equivalent to GLS on (1) as long as there is an intercept in (1). Consider the special case of equicorrelated disturbances Ω = σ 2 [(1 − ρ)IT + ρJT ]

(5)

where IT is an identity matrix of dimension T and JT is a matrix of ones of dimension T . (a) Derive the L and M matrices for the equicorrelated case, and verify the Maeshiro and Wichers result for this special case. (b) Show that for the equicorrelated case, the differencing test given by Plosser, Schwert, and White (1982) can be obtained as the difference between the OLS and GLS estimators of the differenced equation (2). Hint: See the solution by Koning (1992). 18. For the 1982 data on earnings used in Chapter 4, provided as EARN.ASC on the Springer web site compute Ramsey’s (1969) RESET and the Thursby and Schmidt (1977) variant of this test. 19. Repeat problem 18 for the Hedonic housing data given on the Springer web site as HEDONIC.XLS. 20. Repeat problem 18 for the cigarette data given in Table 3.2. 21. Repeat problem 18 for the Gasoline data for Austria given on the Springer web site as GASOLINE.DAT. Use the model given in Chapter 10, section 5. Also compute the PSW differencing test given in (8.54). 22. Use the 1982 data on earnings used in Chapter 4, and provided on the Springer web site as EARN.ASC. Consider the two competing non-nested models H0 ; log(wage) = β 0 + β 1 ED + β 2 EXP + β 3 EXP 2 + β 4 W KS +β 5 M S + β 6 F EM + β 7 BLK + β 8 U N ION + u H1 ; log(wage) = γ 0 + γ 1 ED + γ 2 EXP + γ 3 EXP 2 + γ 4 W KS +γ 5 OCC + γ 6 SOU T H + γ 7 SM SA + γ 8 IN D + ǫ Compute: (a) The Davidson and MacKinnon (1981) J-test for H0 versus H1 . (b) The Fisher and McAleer (1981) JA-test for H0 versus H1 . (c) Reverse the roles of H0 and H1 and repeat parts (a) and (b).


218

Chapter 8: Regression Diagnostics and Specification Tests (d) Both H0 and H1 can be artificially nested in the model used in Chapter 4. Using the F test given in (8.62), test for H0 versus this augmented model. Repeat for H1 versus this augmented model. What do you conclude?

23. For the Consumption-Income data given in Table 5.1, (a) Test the hypothesis that the Consumption model is linear Box-Cox alternative. (b) Test the hypothesis that the Consumption model is log-linear general Box-Cox alternative. 24. Repeat problem 23 for the Cigarette data given in Table 3.2. 25. RESET as a Gauss-Newton Regression. This is based on Baltagi (1998). Davidson and MacKinnon (1993) showed that Ramsey’s (1969) regression error specification test (RESET) can be derived as a Gauss-Newton Regression. This problem is a simple extension of their results. Suppose that the linear regression model under test is given by: yt = Xt′ β + ut

t = 1, 2, . . . , T

(1)

where β is a k × 1 vector of unknown parameters. Suppose that the alternative is the nonlinear regression model between yt and Xt : yt = Xt′ β[1 + θ(Xt′ β) + γ(Xt′ β)2 + λ(Xt′ β)3 ] + ut ,

(2)

where θ, γ, and λ are unknown scalar parameters. It is well known that Ramsey’s (1969) RESET is obtained by regressing yt on Xt , yt2 , yt3 and yt4 and by testing that the coefficients of all powers of yt are jointly zero. Show that this RESET can be derived from a Gauss-Newton Regression on (2), which tests θ = γ = λ = 0.

References This chapter is based on Belsley, Kuh and Welsch (1980), Johnston (1984), Maddala (1992) and Davidson and MacKinnon (1993). Additional references are the following: Baltagi, B.H. (1990), “The Differencing Test in a Regression with Equicorrelated Disturbances,” Econometric Theory, Problem 90.4.5, 6: 488. Baltagi, B.H. (1998), “Regression Specification Error Test as A Gauss-Newton Regression,” Econometric Theory, Problem 98.4.3, 14: 526. Belsley, D.A., E. Kuh and R.E. Welsch (1980), Regression Diagnostics (Wiley: New York). Box, G.E.P. and D.R. Cox (1964), “An Analysis of Transformations,” Journal of the Royal Statistical Society, Series B, 26: 211-252. Brown, R.L., J. Durbin, and J.M. Evans (1975), “Techniques for Testing the Constancy of Regression Relationships Over Time,” Journal of the Royal Statistical Society 37:149-192. Chesher, A. and R. Spady (1991), “Asymptotic Expansions of the Information Matrix Test Statistic,” Econometrica 59: 787-815. Cook, R.D. (1977), “Detection of Influential Observations in Linear Regression,” Technometrics 19:15-18. Cook, R.D. and S. Weisberg (1982), Residuals and Influences in Regression (Chapman and Hall: New York).


References

219

Cox, D.R. (1961), “Tests of Separate Families of Hypotheses,” Proceedings of the Fourth Berkeley Symposium on Mathematical Statistics and Probability, 1: 105-123. Davidson, R., L.G. Godfrey and J.G. MacKinnon (1985), “A Simplified Version of the Differencing Test,” International Economic Review, 26: 639-47. Davidson, R. and J.G. MacKinnon (1981), “Several Tests for Model Specification in the Presence of Alternative Hypotheses,” Econometrica, 49: 781-793. Davidson, R. and J.G. MacKinnon (1985), “Testing Linear and Loglinear Regressions Against Box-Cox Alternatives,” Canadian Journal of Economics, 18: 499-517. Davidson, R. and J.G. MacKinnon (1992), “A New Form of the Information Matrix Test,” Econometrica, 60: 145-157. Davidson, R. and J.G. MacKinnon (2001), “Artificial Regressions,” Chapter 1 in Baltagi, B.H. (ed.) A Companion to Theoretical Econometrics (Blackwell: Massachusetts). Fisher, G.R. and M. McAleer (1981), “Alternative Procedures and Associated Tests of Significance for Non-Nested Hypotheses,” Journal of Econometrics, 16: 103-119. Gentleman, J.F. and M.B. Wilk (1975), “Detecting Outliers II: Supplementing the Direct Analysis of Residuals,” Biometrics, 31: 387-410. Godfrey, L.G. (1988), Misspecification Tests in Econometrics: The Lagrange Multiplier Principle and Other Approaches (Cambridge University Press: Cambridge). Hall, A. (1987), “The Information Matrix Test for the Linear Model,” Review of Economic Studies, 54: 257-263. Harvey, A.C. (1976), “An Alternative Proof and Generalization of a Test for Structural Change,” The American Statistician, 30: 122-123. Harvey, A.C. (1990), The Econometric Analysis of Time Series (MIT Press: Cambridge). Harvey, A.C. and P. Collier (1977), “Testing for Functional Misspecification in Regression Analysis,” Journal of Econometrics, 6: 103-119. Harvey, A.C. and G.D.A. Phillips (1974), “A Comparison of the Power of Some Tests for Heteroskedasticity in the General Linear Model,” Journal of Econometrics, 2: 307-316. Hausman, J. (1978), “Specification Tests in Econometrics,” Econometrica, 46: 1251-1271. Koning, R.H. (1992), “The Differencing Test in a Regression with Equicorrelated Disturbances,” Econometric Theory, Solution 90.4.5, 8: 155-156. Kr¨ amer, W. and H. Sonnberger (1986), The Linear Regression Model Under Test (Physica-Verlag: Heidelberg). Krasker, W.S., E. Kuh and R.E. Welsch (1983), “Estimation for Dirty Data and Flawed Models,” Chapter 11 in Handbook of Econometrics, Vol. I, eds. Z. Griliches and M.D. Intrilligator, Amsterdam, NorthHolland. Maeshiro, A. and R. Wichers (1989), “On the Relationship Between the Estimates of Level Models and Difference Models,” American Journal of Agricultural Economics, 71: 432-434. Orme, C. (1990), “The Small Sample Performance of the Information Matrix Test,” Journal of Econometrics, 46: 309-331. Pagan, A.R. and A.D. Hall (1983), “Diagnostic Tests as Residual Analysis,” Econometric Reviews, 2: 159-254.


220

Chapter 8: Regression Diagnostics and Specification Tests

Pesaran, M.H. and M. Weeks (2001), “Nonnested Hypothesis Testing: A Overview,” Chapter 13 in Baltagi, B.H. (ed.) A Companion to Theoretical Econometrics (Blackwell: Massachusetts). Phillips, G.D.A. and A.C. Harvey (1974), “A Simple Test for Serial Correlation in Regression Analysis,” Journal of the American Statistical Association, 69: 935-939. Plosser, C.I., G.W. Schwert, and H. White (1982), “Differencing as a Test of Specification,” International Economic Review, 23: 535-552. Ramsey, J.B. (1969), “Tests for Specification Errors in Classical Linear Least-Squares Regression Analysis,” Journal of the Royal Statistics Society, Series B, 31: 350-371. Ramsey, J.B. and P. Schmidt (1976), “Some Further Results in the Use of OLS and BLUS Residuals in Error Specification Tests,” Journal of the American Statistical Association, 71: 389-390. Schmidt, P. (1976), Econometrics (Marcel Dekker: New York). Theil, H. (1971), Principles of Econometrics (Wiley: New York). Thursby, J. and P. Schmidt (1977), “Some Properties of Tests for Specification Error in a Linear Regression Model,” Journal of the American Statistical Association, 72: 635-641. Utts, J.M. (1982), “The Rainbow Test for Lack of Fit in Regression,” Communications in Statistics, 11: 2801-2815. Velleman, P. and R. Welsch (1981), “Efficient Computing of Regression Diagnostics,” The American Statistician, 35: 234-242. White, H. (1980), “A Heteroskedasticity–Consistent Covariance Matrix Estimator and a Direct Test for Heteroskedasticity,” Econometrica, 48: 817-838. White, H. (1982), “Maximum Likelihood Estimation of Misspecified Models,” Econometrica, 50: 1-25. Wooldridge, J.M. (1995), “Diagnostic Testing,” Chapter 9 in B.H. Baltagi (ed.) A Companion to Theoretical Econometrics (Blackwell: Massachusetts). Wu, P. (1993), “Variable Addition Test,” Econometric Theory, Problem 93.1.2, 9: 145-146.


CHAPTER 9

Generalized Least Squares 9.1

Introduction

This chapter considers a more general variance covariance matrix for the disturbances. In other words, u ∼ (0, σ 2 In ) is relaxed so that u ∼ (0, σ 2 Ω) where Ω is a positive definite matrix of dimension (n×n). First Ω is assumed known and the BLUE for β is derived. This estimator turns   out to be different from β OLS , and is denoted by β GLS , the Generalized Least Squares estimator  of β. Next, we study the properties of β OLS under this nonspherical form of the disturbances. It turns out that the OLS estimates are still unbiased and consistent, but their standard errors as computed by standard regression packages are biased and inconsistent and lead to misleading inference. Section 9.3 studies some special forms of Ω and derive the corresponding BLUE for β. It turns out that heteroskedasticity and serial correlation studied in Chapter 5 are special cases of Ω. Section 9.4 introduces normality and derives the maximum likelihood estimator. Sections 9.5 and 9.6 study the way in which test of hypotheses and prediction get affected by this general variance-covariance assumption on the disturbances. Section 9.7 studies the properties of this BLUE for β when Ω is unknown, and is replaced by a consistent estimator. Section 9.8 studies what happens to the W, LR and LM statistics when u ∼ N (0, σ 2 Ω). Section 9.9 gives another application of GLS to spatial autocorrelation.

9.2

Generalized Least Squares

The regression equation did not change, only the variance-covariance matrix of the disturbances. It is now σ 2 Ω rather than σ 2 In . However, we can rely once again on a result from matrix algebra to transform our nonspherical disturbances back to spherical form, see the Appendix to Chapter 7. This result states that for every positive definite matrix Ω, there exists a nonsingular matrix P such that P P ′ = Ω. In order to use this result, we transform the original model y = Xβ + u

(9.1)

by premultiplying it by P −1 . We get P −1 y = P −1 Xβ + P −1 u

(9.2)

Defining y ∗ as P −1 y and X ∗ and u∗ similarly, we have y ∗ = X ∗ β + u∗

(9.3)

with u∗ having 0 mean and var(u∗ ) = P −1 var(u)P −1′ = σ 2 P −1 ΩP ′−1 = σ 2 P −1 P P ′ P ′−1 = σ 2 In . Hence, the variance-covariance of the disturbances in (9.3) is a scalar times an identity matrix. Therefore, using the results of Chapter 7, the BLUE for β in (9.1) is OLS on the transformed model in (9.3) ∗′ ∗ −1 ∗′ ∗ ′ −1′ −1  β P X)−1 X ′ P −1′ P −1 y = (X ′ Ω−1 X)−1 X ′ Ω−1 y (9.4) BLU E = (X X ) X y = (X P


222

Chapter 9: Generalized Least Squares

BLU E ) = σ 2 (X ∗′ X ∗ )−1 = σ 2 (X ′ Ω−1 X)−1 . This β  BLU E is known as β  GLS . Define with var(β ′ 2 2 Σ = E(uu ) = σ Ω, then Σ differs from Ω only by the positive scalar σ . One can easily verify ′ −1 −1 ′ −1    that β GLS can be alternatively written as β GLS = (X Σ X) X Σ y and that var(β GLS ) = ′ −1 −1 −1 −1 2  (X Σ X) . Just substitute Σ = Ω /σ in this last expression for β GLS and verify that this yields (9.4).    It is clear that β GLS differs from β OLS . In fact, since β OLS is still a linear unbiased estimator  of β, the Gauss-Markov Theorem states that it must have a variance larger than that of β GLS . ′ −1 ′  Using equation (7.5) from Chapter 7, i.e., β OLS = β + (X X) X u it is easy to show that 2 ′ −1 ′ ′ −1  var(β OLS ) = σ (X X) (X ΩX)(X X)

(9.5)

 GLS )′ Ω−1 (y − X β  GLS )/(n − K) s∗2 = e∗′ e∗ /(n − K) = (y − X β

(9.6)

  Problem 1 shows that var(β OLS )− var(β GLS ) is a positive semi-definite matrix. Note that 2 (X ′ X)−1 , and hence a regression package that is programmed to  var(β ) is no longer σ OLS  compute s2 (X ′ X)−1 as an estimate of the variance of β OLS is using the wrong formula. Fur2 2 thermore, problem 2 shows that E(s ) is not in general σ . Hence, the regression package is also  wrongly estimating σ 2 by s2 . Two wrongs do not make a right, and the estimate of var(β OLS ) is biased. The direction of this bias depends upon the form of Ω and the X matrix. (We saw some examples of this bias under heteroskedasticity and serial correlation in Chapter 5). Hence, the standard errors and t-statistics computed using this OLS regression are biased. Under heteroskedasticity, one can use the White (1980) robust standard errors for OLS. In this case,  = diag[e2 ] where ei denotes the least squares residuals. Σ = σ 2 Ω in (9.5) is estimated by Σ i The resulting t-statistics are robust to heteroskedasticity. Similarly Wald type statistics for 2 ′ −1 in (7.41) by (9.5) with  H0 ; Rβ = r can be obtained based on β OLS by replacing σ (X X) 2  Σ = diag[ei ]. In the presence of both serial correlation and heteroskedasticity, one can use the consistent covariance matrix estimate suggested by Newey and West (1987). This was discussed in Chapter 5.  To summarize, β OLS is no longer BLUE whenever Ω = In . However, it is still unbiased and consistent. The last two properties do not rely upon the form of the variance-covariance matrix of the disturbances but rather on E(u/X) = 0 and plim X ′ u/n = 0. The standard errors of  β OLS as computed by the regression package are biased and any test of hypothesis based on this OLS regression may be misleading. So far we have not derived an estimator for σ 2 . We know however, from the results in Chapter 7, that the transformed regression (9.3) yields a mean squared error that is an unbiased estimator for σ 2 . Denote this by s∗2 which is equal to the transformed OLS residual sum of squares divided by (n − K). Let e∗ denote the vector of OLS residuals from (9.3), this means that −1 (y − X β −1 e   e∗ = y ∗ − X ∗ β GLS and GLS = P GLS ) = P = e′GLS Ω−1 eGLS /(n − K)

Note that s∗2 now depends upon Ω−1 .

Necessary and Sufficient Conditions for OLS to be Equivalent to GLS There are several necessary and sufficient conditions for OLS to be equivalent to GLS, see Puntanen and Styan (1989) for a historical survey. For pedagogical reasons, we focus on the derivation of Milliken and Albohali (1984). Note that y = PX y + P¯X y. Therefore, replacing y


9.3

Special Forms of Ω

223

GLS by this expression we get in β

′ −1 −1 ′ −1 ′ −1 −1 ′ −1 ¯   ¯ β GLS = (X Ω X) X Ω [PX y + PX y] = β OLS + (X Ω X) X Ω PX y

The last term is zero for every y if and only if X ′ Ω−1 P¯X = 0

(9.7)

  Therefore, β GLS = β OLS if and only if (9.7) is true. Another easy necessary and sufficient condition to check in practice is the following: PX Ω = ΩPX

(9.8)

see Zyskind (1967). This involves Ω rather than Ω−1 . There are several applications in economics where these conditions are satisfied and can be easily verified, see Balestra (1970) and Baltagi (1989). We will apply these conditions in Chapter 10 on Seemingly Unrelated Regressions, Chapter 11 on simultaneous equations and Chapter 12 on panel data. See also problem 9.

9.3

Special Forms of Ω

If the disturbances are heteroskedastic but not serially correlated, then Ω = diag[σ 2i ]. In this case, P = diag[σ i ], P −1 = Ω−1/2 = diag[1/σ i ] and Ω−1 = diag[1/σ 2i ]. Premultiplying the regression equation by Ω−1/2 is equivalent to dividing the i-th observation of this model by σ i . This makes the new disturbance ui /σ i have 0 mean and homoskedastic variance σ 2 , leaving properties ∗ = X /σ for like no serial correlation intact. The new regression runs yi∗ = yi /σ i on Xik i ik i = 1, 2, . . . , n, and k = 1, 2, . . . , K. Specific assumptions on the form of these σ i ’s were studied in the heteroskedasticity chapter. If the disturbances follow an AR(1) process ut = ρut−1 + ǫt for t = 1, 2, . . . , T ; with |ρ| ≤ 1 and ǫt ∼IID(0, σ 2ǫ ), then cov(ut , ut−s ) = ρs σ 2u with σ 2u = σ 2ǫ /(1 − ρ2 ). This means that ⎡ ⎤ 1 ρ ρ2 . . . ρT −1 ⎢ ρ 1 ρ . . . ρT −2 ⎥ ⎥ Ω=⎢ (9.9) ⎣ : : : : ⎦ T −1 T −2 T −3 ρ ρ ρ ... 1 and

Ω−1 =

Then

P −1



1 1 − ρ2

⎡ 

⎢ ⎢ ⎢ =⎢ ⎢ ⎢ ⎣

⎢ ⎢ ⎢ ⎢ ⎣

1 −ρ 0 ... 0 0 0 −ρ 1 + ρ2 −ρ . . . 0 0 0 : : : : : : 0 0 0 . . . −ρ 1 + ρ2 −ρ 0 0 0 ... 0 −ρ 1

1 − ρ2 0 0 . . . 0 0 −ρ 1 0 ... 0 0 0 −ρ 1 . . . 0 0 : : : : : 0 0 0 . . . −ρ 1 0 0 0 . . . 0 −ρ

0 0 0 : 0 1

⎤ ⎥ ⎥ ⎥ ⎥ ⎦

(9.10)

⎤ ⎥ ⎥ ⎥ ⎥ ⎥ ⎥ ⎦

(9.11)


224

Chapter 9: Generalized Least Squares

is the matrix that satisfies the following condition P −1′ P −1 = (1 − ρ2 )Ω−1 . Premultiplying the regression model by P −1 is equivalent to performing the Prais-Winsten transformation. In  particular the first observation on y becomes y1∗ = 1 − ρ2 y1 and the remaining observations are given by yt∗ = (yt −ρyt−1 ) for t = 2, 3, . . . , T , with similar terms for the X’s and the disturbances. Problem 3 shows that the variance covariance matrix of the transformed disturbances u∗ = P −1 u is σ 2ǫ IT . Other examples where an explicit form for P −1 has been derived include, (i) the MA(1) model, see Balestra (1980); (ii) the AR(2) model, see Lempers and Kloek (1973); (iii) the specialized AR(4) model for quarterly data, see Thomas and Wallis (1971); and (iv) the error components model, see Fuller and Battese (1974) and Chapter 12.

9.4

Maximum Likelihood Estimation

Assuming that u ∼ N (0, σ 2 Ω), the new likelihood function can be derived keeping in mind that u∗ = P −1 u = Ω−1/2 u and u∗ ∼ N (0, σ 2 In ). In this case f (u∗1 , . . . , u∗n ; β, σ 2 ) = (1/2πσ 2 )n/2 exp{−u∗′ u∗ /2σ 2 }

(9.12)

Making the transformation u = P u∗ = Ω1/2 u∗ , we get f (u1 , . . . , un ; β, σ 2 ) = (1/2πσ 2 )n/2 |Ω−1/2 | exp{−u′ Ω−1 u/2σ 2 }

(9.13)

where |Ω−1/2 | is the Jacobian of the inverse transformation. Finally, substituting y = Xβ + u in (9.13), one gets the likelihood function L(β, σ 2 ; Ω) = (1/2πσ 2 )n/2 |Ω−1/2 | exp{−(y − Xβ)′ Ω−1 (y − Xβ)/2σ 2 }

(9.14)

since the Jacobian of this last transformation is 1. Knowing Ω, maximizing (9.14) with respect to  β is equivalent to minimizing u∗′ u∗ with respect to β. This means that β M LE is the OLS estimate  on the transformed model, i.e., β . From (9.14), we see that this RSS is a weighted one with GLS the weight being the inverse of the variance covariance matrix of the disturbances. Similarly, maximizing (9.14) with respect to σ 2 gets σ 2M LE = the OLS residual sum of squares of the transformed regression (9.3) divided by n. From (9.6) this can be written as σ 2M LE = e∗′ e∗ /n = (n − K)s∗2 /n. The distributions of these maximum likelihood estimates can be derived from 2 ′ −1 −1  the transformed model using the results in Chapter 7. In fact, β GLS ∼ N (β, σ (X Ω X) ) ∗2 2 2 and (n − K)s /σ ∼ χn−K .

9.5

Test of Hypotheses

In order to test H0 ; Rβ = r, under the general variance-covariance matrix assumption, one can revert to the transformed model (9.3) which has a scalar identity variance-covariance matrix and use the test statistic derived in Chapter 7 GLS − r)/σ 2 ∼ χ2  GLS − r)′ [R(X ∗′ X ∗ )−1 R′ ]−1 (Rβ (Rβ g

(9.15)

′ ′ −1 −1 ′ −1 2 2   (Rβ GLS − r) [R(X Ω X) R ] (Rβ GLS − r)/σ ∼ χg

(9.16)

GLS replaces β OLS and X ∗ replaces X. Replacing X ∗ by P −1 X, we get Note that β


9.6

Prediction

225

GLS replaces This differs from its counterpart in the spherical disturbances model in two ways. β ′ −1 ′  β OLS , and (X Ω X) takes the place of X X. One can also derive the restricted estimator based on the transformed model by simply replacing X ∗ by P −1 X and the OLS estimator of β by its GLS counterpart. Problem 4 asks the reader to verify that the restricted GLS estimator is ′ −1 −1 ′ ′ −1 −1 ′ −1    β RGLS = β GLS − (X Ω X) R [R(X Ω X) R ] (Rβ GLS − r)

(9.17)

Furthermore, using the same analysis given in Chapter 7, one can show that (9.15) is in fact the Likelihood Ratio statistic and is equal to the Wald and Lagrangian Multiplier statistics, see Buse (1982). In order to operationalize these tests, we replace σ 2 by its unbiased estimate s∗2 , and divide by g the number of restrictions. The resulting statistic is an F (g, n − K) for the same reasons given in Chapter 7.

9.6

Prediction

How is prediction affected by nonspherical disturbances? Suppose we want to predict one period ahead. What has changed with a general Ω? For one thing, we now know that the period (T + 1) disturbance is correlated with the sample disturbances. Let us assume that this correlation is given by the (T × 1) vector ω = E(uT +1 u), problem 5 shows that the BLUP for yT +1 is  GLS + ω′ Ω−1 (y − X β  GLS )/σ 2 yT +1 = x′T +1 β

(9.18)

The first term is as expected, however it is the second term that highlights the difference between the spherical and nonspherical model predictions. To illustrate this, let us look at the AR(1) case where cov(ut , ut−s ) = ρs σ 2u . This implies that ω ′ = σ 2u (ρT , ρT −1 , . . . , ρ). Using Ω which is given in (9.9), one can show that ω is equal to ρσ 2u multiplied by the last column of Ω. But Ω−1 Ω = IT , therefore, Ω−1 times the last column of Ω gives the last column of the identity matrix, i.e., (0, 0, . . . , 1)′ . Substituting for the last column of Ω its expression (ω/ρσ 2u ) one gets, Ω−1 (ω/ρσ 2u ) = (0, 0, . . . , 1)′ . Transposing and rearranging this last expression, we get ω ′ Ω−1 /σ 2u = ρ(0, 0, . . . , 1). This means that the last term in (9.18) is equal to ρ(0, 0, . . . , 1)(y −  Xβ GLS ) = ρeT,GLS , where eT,GLS is the T -th GLS residual. This differs from the spherical model prediction in that next year’s disturbance is not independent of the sample disturbances and hence, is not predicted by its mean which is zero. Instead, one uses the fact that uT +1 = ρuT + ǫT +1 and predicts uT +1 by ρeT,GLS . Only ǫT +1 is predicted by its zero mean but uT is predicted by eT,GLS .

9.7

Unknown Ω

 and substitute that in If Ω is unknown, the practice is to get a consistent estimate of Ω, say Ω  β GLS . The resulting estimator is −1 ′  −1 ′  −1  β F GLS = (X Ω X) X Ω y

(9.19)

 replaces Ω the Gauss-Markov Theorem and is called a feasible GLS estimator of β. Once Ω  no longer necessarily holds. In other words, β F GLS is not BLUE, although it is still consistent.


226

Chapter 9: Generalized Least Squares

 F GLS are in general difficult to derive. However, we have the The finite sample properties of β following asymptotic results. √  √  n(β F GLS − β) have the same asymptotic distribution Theorem 1: n(β GLS − β) and  −1 − Ω−1 )X/n = 0 N (0, σ 2 Q−1 ), where Q = lim(X ′ Ω−1 X)/n as n → ∞, if (i) plim X ′ (Ω ′ −1 −1   and (ii) plim X (Ω − Ω )u/n = 0. A sufficient condition for this theorem to hold is that Ω is a consistent estimator of Ω and X has a satisfactory limiting behavior.

 −1 − Ω−1 )u/n = 0, then s∗2 = e′ Ω−1 eGLS /(n − K) and Lemma 1: If in addition plim u′ (Ω GLS  −1 eF GLS /(n − K) are both consistent for σ 2 . This means that one can perform s∗2 = e′F GLS Ω  F GLS and s∗2 rather than β GLS and test of hypotheses based on asymptotic arguments using β ∗2 s , respectively. For a proof of Theorem 1 and Lemma 1, see Theil (1971), Schmidt (1976) or Judge et al. (1985). Monte Carlo evidence under heteroskedasticity or serial correlation suggest that there is gain in performing feasible GLS rather than OLS in finite samples. However, we have also seen in Chapter 5 that performing a two-step Cochrane-Orcutt procedure is not necessarily better than OLS if the X’s are trended. This says that feasible GLS omitting the first observation (in this case Cochrane-Orcutt) may not be better in finite samples than OLS using all the observations.

9.8

The W, LR and LM Statistics Revisited

In this section we present a simplified and more general proof of W ≥ LR ≥ LM due to Breusch (1979). For the general linear model given in (9.1) with u ∼ N (0, Σ) and H0 ; Rβ = r. The likelihood function given in (9.14) with Σ = σ 2 Ω, can be maximized with respect to  and Σ,  =  where β β and Σ without imposing H0 , yielding the unrestricted estimators β u u ′ −1 −1 ′ −1   (X Σ X) X Σ y. Similarly, this likelihood can be maximized subject to the restriction H0 ,  and Σ,  where yielding β r r = (X ′ Σ  −1 X)−1 X ′ Σ  −1 y − (X ′ Σ  −1 X)−1 R′ μ  β

(9.20)

 = (X ′ Σ  −1 X)−1 X ′ Σ  −1 y β r

(9.21)

 − (X ′ Σ  − r)  =β  −1 X)−1 R′ A −1 (Rβ β u u u

(9.22)

 Σ)/L(β,  Σ)] LR = −2log[max L(β/Σ)/max L(β/Σ)] = −2log[L(β,

(9.23)

 − r) is the Lagrange multiplier described in equation (7.35) of −1 (Rβ as in (9.17), where μ =A r ′ −1   Chapter 7 and A = [R(X Σ X)−1 R′ ]. The major distinction from Chapter 7 is that Σ is un denote the unrestricted maximum likelihood estimator of known and has to be estimated. Let β r  denote the restricted  and let β β conditional on the restricted variance-covariance estimator Σ u maximum likelihood of β (satisfying H0 ) conditional on the unrestricted variance-covariance  More explicitly, estimator Σ. and

Knowing Σ, the Likelihood Ratio statistic is given by Rβ=r

= u ′ Σ−1 u −u ′ Σ−1 u 

β


Spatial Error Correlation

227

 and u  both estimators of β are conditional on a known Σ. where u  = y − Xβ  = y − X β,  ∼ N (Rβ, R(X ′ Σ  −1 X)−1 R′ ) Rβ u

and the Wald statistic is given by u − r) u − r)′ A −1 (Rβ W = (Rβ

 = [R(X ′ Σ  −1 X)−1 R′ ] where A (9.24)  and u  are related as follows: Using (9.22), it is easy to show that u u = y − X β u = y − X β u u −1 ′ −1 ′  −1  (9.25) u u = u u + X(X Σ X) R A (Rβ u − r) and

 − r)  − r)′ A −1 (Rβ  −1 u  −1 u u + (Rβ u = u ′u Σ u ′u Σ u u ′ −1  The cross-product terms are zero because X Σ u u = 0. Therefore, ′  −1 ′  −1   Σ)]   u = −2log[L(β, Σ)/L( β, u − u  Σ u W = u  Σ u u

u

(9.26)

(9.27)

  L(β/Σ)] = −2log[max L(β/Σ)/max β

Rβ=r

 the unrestricted and the Wald statistic can be interpreted as a LR statistic conditional on Σ, maximum likelihood estimator of Σ. Similarly, the Lagrange multiplier statistic, which tests that μ = 0, is given by  − r)  − r)′ A −1 (Rβ  = (Rβ (9.28) LM = μ′ Aμ r r

Using (9.20) one can easily show that  − r)  −1 X)−1 R′ A −1 (Rβ u r = u r + X(X ′ Σ r

(9.29)

and

 −1 u   −1 u r + μ′ Aμ r = u ′r Σ u ′r Σ

 −1 u X ′Σ r

The cross-product terms are zero because = 0. Therefore, ′  −1 ′  −1  , Σ)/L(  , Σ)]   LM = u r Σ u r − u r Σ u r = −2log[L(β β r r

(9.30) (9.31)

  = −2log[max L(β/Σ)/max L(β/Σ)] β

Rβ=r

 the and the Lagrange multiplier statistic can be interpreted as a LR statistic conditional on Σ restricted maximum likelihood of Σ. Given that  ≤ max L(β, Σ) = max L(β/Σ)  (9.32) max L(β/Σ) β

β,Σ

β

 ≤ max L(β, Σ) = max L(β/Σ)  max L(β/Σ)

(9.33)

LR = −2log[max L(β, Σ)/max L(β, Σ)]

(9.34)

Rβ=r

Rβ=r,Σ

Rβ=r

it can be easily shown that the likelihood ratio statistic given by Rβ=r,Σ

β,Σ

satisfies the following inequality W ≥ LR ≥ LM

(9.35)

The proof is left to the reader, see problem 6. This general and simple proof holds as long as the maximum likelihood estimator of β is uncorrelated with the maximum likelihood estimator of Σ, see Breusch (1979).


228

Chapter 9: Generalized Least Squares

9.9

Spatial Error Correlation1

Unlike time-series, there is typically no unique natural ordering for cross-sectional data. Spatial autocorrelation permit correlation of the disturbance terms across cross-sectional units. There is an extensive literature on spatial models in regional science, urban economics, geography and statistics, see Anselin (1988). Examples in economics usually involve spillover effects or externalities due to geographical proximity. For example, the productivity of public capital, like roads and highways, on the output of neighboring states. Also, the pricing of welfare in one state that pushes recipients to other states. Spatial correlation could relate directly to the model dependent variable y, the exogenous variables X, the disturbance term u, or to a combination of all three. Here we consider spatial correlation in the disturbances and leave the remaining literature on spatial dependence to the motivated reader to pursue in Anselin (1988, 2001) and Anselin and Bera (1998) to mention a few. For the cross-sectional disturbances, the spatial autocorrelation is specified as u = λW u + ǫ

(9.36)

where λ is the spatial autoregressive coefficient satisfying |λ| < 1, and ǫ ∼ IIN(0, σ 2 ). W is a known spatial weight matrix with diagonal elements equal to zero. W also satisfies some other regularity conditions like the fact that In − λW must be nonsingular. The regression model given in (9.1) can be written as y = Xβ + (In − λW )−1 ǫ

(9.37)

with the variance-covariance matrix of the disturbances given by Σ = σ 2 Ω = σ 2 (In − λW )−1 (In − λW ′ )−1

(9.38)

Under normality of the disturbances, Ord (1975) derived the maximum likelihood estimators n 1 lnL = − ln|Ω| − ln2πσ 2 − (y − Xβ)′ Ω−1 (y − Xβ)/2σ 2 2 2

(9.39)

The Jacobian term simplifies by using ln|Ω| = −2ln|I − λW | = −2

n

i=1 ln(1

− λwi )

(9.40)

where wi are the eigenvalues of the spatial weight matrix W . The first-order conditions yield the familiar GLS estimator of β and the associated estimator of σ 2 : ′ −1 −1 ′ −1  β M LE = (X Ω X) X Ω y

and

σ 2M LE = e′M LE Ω−1 eM LE /n

(9.41)

 where eM LE = y − X β M LE . An estimate of λ can be obtained using the iterative solution of the first-order conditions in Magnus (1978, p. 283): 1 − tr 2



∂Ω−1 ∂λ

  −1  ∂Ω ′ Ω = eM LE eM LE ∂λ

(9.42)


Note

229

where ∂Ω−1 /∂λ = −W − W ′ + λW ′ W

(9.43)

 Alternatively, one can substitute β 2M LE from (9.41) into the log-likelihood in (9.39) M LE and σ to get the concentrated log-likelihood which will be a nonlinear function of λ, see Anselin (1988) for details. Testing for zero spatial autocorrelation i.e., H0 ; λ = 0 is usually based on the Moran I-test which is similar to the Durbin-Watson statistic in time-series. This is given by   n e′ W e (9.44) MI = S0 e′ e where e denotes the vector OLS residuals and S0 is a standardization factor equal to the sum  of of the spatial weights ni=1 nj=1 wij . For a row-standardized weights matrix W where each row sums to one, S0 = n and the Moran I-statistic simplifies to e′ W e/e′ e. In practice the test is implemented by standardizing it and using the asymptotic N (0, 1) critical values, see Anselin and Bera (1988). In fact, for a row-standardized W matrix, the mean and variance of the Moran I-statistic is obtained from   ′ e We = tr(P¯X W )/(n − k) (9.45) E(M I) = E e′ e and E(M I)2 =

tr(P¯X W P¯X W ′ ) + tr(P¯X W )2 + {tr(P¯X¯ W )}2 (n − k)(n − k + 2)

Alternatively, one can derive the Lagrange Multiplier test for H0 ; λ = 0 using the result that ∂lnL/∂λ evaluated under the null of λ = 0 is equal to u′ W u/σ 2 and the fact that the Information matrix is block-diagonal between β and (σ 2 , λ), see problem 14. In fact, one can show that LMλ =

(e′ W e/ σ 2 )2 tr[(W ′ + W )W ]

(9.46)

with σ 2 = e′ e/n. Under H0 , LMλ is asymptotically distributed as χ21 . One can clearly see the connection between Moran’s I-statistic and LMλ . Computationally, the W and LR tests are more demanding since the require ML estimation under spatial autocorrelation. This is only a brief introduction into the spatial dependence literature. Hopefully, it will motivate the reader to explore alternative formulations of spatial dependence, alternative estimation and testing methods discussed in this literature and the numerous applications in economics on hedonic housing, crime rates, police expenditures and R&D spillovers, to mention a few.

Note 1. This section is based on Anselin (1988, 2001) and Anselin and Bera (1998).


230

Chapter 9: Generalized Least Squares

Problems 1.

 (a) Using equation (7.5) of Chapter 7, verify that var(β OLS ) is that given in (9.5). 2 ′   (b) Show that var(β OLS )− var(β GLS ) = σ AΩA where A = [(X ′ X)−1 X ′ − (X ′ Ω−1 X)−1 X ′ Ω−1 ].

Conclude that this difference in variances is positive semi-definite. 2.

(a) Show that E(s2 ) = σ 2 tr(ΩP¯X )/(n − K) = σ 2 . Hint: Follow the same proof given below equation (7.6) of Chapter 7, but substitute σ 2 Ω instead of σ 2 In . (b) Use the fact that PX and Σ are non-negative definite with tr(ΣPX ) ≥ 0 to show nmatrices 2 that 0 ≤ E(s2 ) ≤ tr(Σ)/(n − K) where tr(Σ) = σ with σ 2i = var(ui ) ≥ 0. This i=1 i bound was derived by Dufour (1986). Under homoskedasticity, show that this bound becomes 0 ≤ E(s2 ) ≤ nσ 2 /(n − K). In general, 0 ≤ {mean of n − K smallest characteristic roots of Σ} ≤ E(s2 ) ≤ {mean of n − K largest characteristic roots of Σ} ≤ tr(Σ)/(n − K), see Sathe and Vinod (1974) and Neudecker (1977, 1978). (c) Show that a sufficient condition for s2 to be consistent for σ 2 irrespective of X is that λmax = the largest characteristic root of Ω is o(n), i.e., λmax /n → 0 as n → ∞ and plim (u′ u/n) = σ 2 . Hint: s2 = u′ P¯X u/(n − K) = u′ u/(n − K) − u′ PX u/(n − K). By assumption, the first term tends in probability limits to σ 2 as n → ∞. The second term has expectation σ 2 tr(PX Ω)/(n − K). Now PX Ω has rank K and therefore exactly K non-zero characteristic roots each of which cannot exceed λmax . This means that E[u′ PX u/(n−K)] ≤ σ 2 Kλmax /(n− K). Using the condition that λmax /n → 0 proves the result. See Kr¨amer and Berghoff (1991).

(d) Using the same reasoning in part (a), show that s∗2 given in (9.6) is unbiased for σ 2 . 3. The AR(1) Model. See Kadiyala (1968). (a) Verify that ΩΩ−1 = IT for Ω and Ω−1 given in (9.9) and (9.10), respectively. (b) Show that P −1′ P −1 = (1 − ρ2 )Ω−1 for P −1 defined in (9.11).

(c) Conclude that var(P −1 u) = σ 2ǫ IT . Hint: Ω = (1 − ρ2 )P P ′ as can be easily derived from part (b).

4. Restricted GLS. Using the derivation of the restricted least squares estimator for u ∼ (0, σ 2 In ) in Chapter 7, verify equation (9.17) for the restricted GLS estimator based on u ∼ (0, σ 2 Ω). Hint: Apply restricted least squares results to the transformed model given in (9.3). 5. Best Linear Unbiased Prediction. This is based on Goldberger (1962). Consider all linear predictors of yT +s = x′T +s β + uT +s of the form yT +s = c′ y, where u ∼ (0, Σ) and Σ = σ 2 Ω. (a) Show that c′ X = x′T +s for yT +s to be unbiased.

(b) Show that var( yT +s ) = c′ Σc + σ 2T +s − 2c′ ω where var(uT +s ) = σ 2T +s and ω = E(uT +s u). (c) Minimize var( yT +s ) given in part (b) subject to c′ X = x′T +s and show that  c = Σ−1 [IT − X(X ′ Σ−1 X)−1 X ′ Σ−1 ]ω + Σ−1 X(X ′ Σ−1 X)−1 xT +s

′ −1 ′ −1   This means that yT +s =  c′ y = x′T +s β eGLS = x′T +s β eGLS /σ 2 . For GLS + ω Σ GLS + ω Ω s = 1, i.e., predicting one period ahead, this verifies equation (9.18). Hint: Use partitioned inverse in solving the first-order minimization equations. s  (d) Show that yT +s = x′T +s β GLS + ρ eT,GLS for the stationary AR(1) disturbances with autoregressive parameter ρ, and |ρ| < 1.


Problems

231

6. The W, LR and LM Inequality. Using the inequalities given in equations (9.32) and (9.33) verify equation (9.35) which states that W ≥ LR ≥ LM . Hint: Use the conditional likelihood ratio interpretations of W and LM given in equations (9.27) and (9.31) respectively. 7. Consider the simple linear regression yi = α + βXi + ui

i = 1, 2, . . . , n

with ui ∼ IIN(0, σ 2 ). For H0 ; β = 0, derive the LR, W and LM statistics in terms of conditional likeσ 2 )/ lihood ratios as described in Breusch (1979). In other words, compute W = −2 log[max (α, β/ H0

max L(α, β/ σ 2 )/max L(α, β/ σ 2 )] and LR = −2log[max L(α, β, σ2 )/ σ 2 )], LM = −2log[max L(α, β/ H0

α,β

α,β

H0

(α, β, σ2 )] where σ  2 is the unrestricted MLE of σ 2 while σ 2 is the restricted MLE of σ 2 under max L 2 α,β,σ

H0 . Use these results to infer that W ≥ LR ≥ LM .

8. Sampling Distributions and Efficiency Comparison of OLS and GLS. Consider the following regression model yt = βxt + ut for (t = 1, 2), where β = 2 and xt takes on the fixed values x1 = 1, x2 = 2. The ut ’s have the following discrete joint probability distribution: (u1 , u2 ) (−1, −2) (1, −2) (−1, 2) (1, 2)

Probability 1/8 3/8 3/8 1/8

(a) What is the variance-covariance matrix of the disturbances? Are the disturbances heteroskedastic? Are they correlated?     (b) Find the sampling distributions of β OLS and β GLS and verify that var(β OLS ) > var(β GLS ).  (c) Find the sampling distribution of the OLS residuals and verify that the estimated var(β OLS ) is biased. Also, find the sampling distribution of the GLS residuals and verify that the MSE of the GLS regression is an unbiased estimator of the GLS regression variance. Hint: Read Oksanen (1991) and Phillips and Wickens (1978), pp. 3-4. This problem is based on Baltagi (1992). See also the solution by Im and Snow (1993). 9. Equi-correlation. This problem is based on Baltagi (1998). Consider the regression model given in (9.1) with equi-correlated disturbances, i.e., equal variances and equal covariances: E(u′ u) = σ 2 Ω = σ 2 [(1 − ρ)IT + ριT ι′T ] where ιT is a vector of ones of dimension T and IT is the identity matrix. In this case, var(ut ) = σ 2 and cov(ut , us ) = ρσ 2 for t = s with t = 1, 2, . . . , T . Assume that the regression has a constant. (a) Show that OLS on this model is equivalent to GLS. Hint: Verify Zyskind’s condition given in (9.8) using the fact that PX ιT = ιT if ιT is a column of X. (b) Show that E(s2 ) = σ 2 (1 − ρ). Also, that Ω is positive semi-definite when −1/(T − 1) ≤ ρ ≤ 1. Conclude that if −1/(T − 1) ≤ ρ ≤ 1, then 0 ≤ E(s2 ) ≤ [T /(T − 1)]σ 2 . The lower and upper bounds are attained at ρ = 1 and ρ = −1/(T − 1), respectively, see Dufour (1986). Hint: Ω is positive semi-definite if for every arbitrary non-zero vector a we have a′ Ωa ≥ 0. What is this expression for a = ιT ? (c) Show that for this equi-correlated regression model, the BLUP of yT +1 = x′T +1 β + uT +1 is  yT +1 = x′T +1 β OLS as long as there is a constant in the model.

10. Consider the simple regression with no regressors and equi-correlated disturbances: yi = α + u i

i = 1, . . . , n


232

Chapter 9: Generalized Least Squares where E(ui ) = 0 and cov(ui , uj )

with

1 (n−1)

= ρσ 2 for i = j = σ 2 for i = j

≤ ρ ≤ 1 for the variance-covariance matrix of the disturbances to be positive definite.

(a) Show that the OLS and GLS estimates of α are identical. This is based on Kruskal (1968). (b) Show that the bias in s2 , the OLS estimator of σ 2 , is given by −ρσ 2 . (c) Show that the GLS estimator of σ 2 is unbiased.

(d) Show that the E[estimated var( α)− true var( αOLS )] is also −ρσ 2 . 11. Prediction Error Variances Under Heteroskedasticity. This is based on Termayne (1985). Consider the t-th observation of the linear regression model given in (9.1). yt = x′t β + ut

t = 1, 2, . . . , T

where yt is a scalar x′t is 1 × K and β is a K × 1 vector of unknown coefficients. ut is assumed to have zero mean, heteroskedastic variances E(u2t ) = (zt′ γ)2 where zt′ is a 1 × r vector of observed variables and γ is an r × 1 vector of parameters. Furthermore, these ut ’s are not serially correlated, so that E(ut us ) = 0 for t = s.   (a) Find the var(β OLS ) and var(β GLS ) for this model.

(b) Suppose we are forecasting y for period f in the future knowing xf , i.e., yf = x′f β + uf with f > T . Let ef and  ef be the forecast errors derived using OLS and GLS, respectively. Show that the prediction error variances of the point predictions of yf are given by T T T var( ef ) = x′f ( t=1 xt x′t )−1 [ t=1 xt x′t (zt′ γ)2 ]( t=1 xt x′t )−1 xf + (zf′ γ)2 var( ef ) =

T x′f [ t=1 xt x′t (zt′ γ)2 ]−1 xf + (zf′ γ)2

(c) Show that the variances of the two forecast errors of conditional mean E(yf /xf ) based   upon β cf and  cf , respectively are the first two terms of the OLS and β GLS and denoted by  corresponding expressions in part (b).

(d) Now assume that K = 1 and r = 1 so that there is only one single regressor xt and one zt variable determining the heteroskedasticity. Assume also for simplicity that the empirical moments of xt match the population moments of a Normal random variable with mean zero and variance θ. Show that the relative efficiency of the OLS to the GLS predictor of yf is equal to (T + 1)/(T + 3), whereas the relative efficiency of the corresponding ratio involving the two predictions of the conditional mean is (1/3). 12. Estimation of Time Series Regressions with Autoregressive Disturbances and Missing Observations. This is based on Baltagi and Wu (1997). Consider the following time series regression model, yt = x′t β + ut

t = 1, . . . , T,

where β is a K × 1 vector of regression coefficients including the intercept. The disturbances follow a stationary AR(1) process, that is, ut = ρut−1 + ǫt ,


Problems

233

with |ρ| < 1, ǫt is IIN(0, σ 2ǫ ), and u0 ∼ N (0, σ 2ǫ /(1 − ρ2 )). This model is only observed at times tj for j = 1, . . . , n with 1 = t1 < . . . < tn = T and n > K. The typical covariance element of ut for the observed periods tj and ts is given by cov(utj , uts ) =

σ 2ǫ ρ|tj −ts | 1 − ρ2

for

s, j = 1, . . . , n

Knowing ρ, derive a simple Prais-Winsten-type transformation that will obtain GLS as a simple least squares regression. 13. Multiplicative Heteroskedasticity. This is based on Harvey (1976). Consider the linear model given in (9.1) and let u ∼ N (0, Σ) where Σ = diag[σ2i ]. Assume that σ 2i = σ 2 hi (θ) with θ ′ = (θ1 , . . . , θs ) and hi (θ) = exp(θ 1 z1i + . . . + θs zsi ) = exp(zi′ θ) with zi′ = (z1i , . . . , zsi ). (a) Show that log-likelihood function is given by log L(β, θ, σ2 ) = −

N 1 N (yi − x′i β)2 1 N log 2πσ 2 − i=1 log hi (θ) − 2 2 2σ 2 i=1 hi (θ)

and the score with respect to θ is ∂log L/∂θ = −

1 N (yi − x′i β)2 ∂hi 1 ∂hi 1 N + 2 i=1 · i=1 2 hi (θ) ∂θ 2σ (hi (θ))2 ∂θ

Conclude that for multiplicative heteroskedasticity, equating this score to zero yields N

i=1

N (yi − x′i β)2 zi = σ 2 i=1 zi . exp(zi′ θ)

(b) Show that the Information matrix is given by ⎡ ′ −1 0 XΣ X ⎢ ⎢ 1 ∂hi ∂hi 1 N ⎢ 0 ⎢ 2 I(β, θ, σ ) = ⎢ 2 i=1 (hi (θ))2 ∂θ ∂θ ′ ⎢ ⎢ ⎣ 1 ∂hi 1 N 0 2σ 2 i=1 hi (θ) ∂θ ′

and for multiplicative heteroskedasticity this becomes ⎡ ′ −1 XΣ X 0 0 ⎢ ⎢ 1 N 1 ′ ⎢ 2 ZZ zi 0 I(β, θ, σ ) = ⎢ 2 2σ 2 i=1 ⎢ ⎢ ⎣ N 1 N ′ z 0 2σ 2 i=1 i 2σ 4

where Zi′ = (z1 , . . . , zN ).

0 1 N 1 ∂hi 2σ 2 i=1 hi (θ) ∂θ N 2σ 4

⎤ ⎥ ⎥ ⎥ ⎥ ⎥ ⎥ ⎥ ⎦

⎤ ⎥ ⎥ ⎥ ⎥ ⎥ ⎥ ⎦

(c) Assume that hi (θ) satisfies hi (0) = 1, then the test for heteroskedasticity is H0 ; θ = 0 versus H1 ; θ = 0. Show that the score with respect to θ and σ 2 evaluated under the null hypothesis, i.e., at θ = 0 and σ  2 = e′ e/N is given by ⎛  ⎞  2 1 N ei − 1 z i ⎠ S = ⎝ 2 i=1 σ 2 0


234

Chapter 9: Generalized Least Squares where e denotes the vector of OLS residuals. The Information matrix with respect to θ and Ď&#x192; 2 can be obtained from the bottom right block of I(β, θ, Ď&#x192; 2 ) given in part (b). Conclude that the score test for H0 is given by

LM =

N

â&#x20AC;˛ 2 i=1 zi (ei

â&#x2C6;&#x2019;Ď&#x192; 2 )



N i=1 (zi

â&#x20AC;˛

â&#x2C6;&#x2019; zÂŻ) (zi â&#x2C6;&#x2019; zÂŻ) 2 Ď&#x192;4

â&#x2C6;&#x2019;1 

N 2 i=1 zi (ei

â&#x2C6;&#x2019;Ď&#x192; 2 )

This statistic is asymptotically distributed as Ď&#x2021;2s under H0 . From Chapter 5, we can see that this is a special case of the Breusch and Pagan (1979) test-statistic which can be obtained as one-half the regression sum of squares of e2 / Ď&#x192; 2 on a constant and Z. Koenker and Bassett  2 2 )2 /N to make this test more (1982) suggested replacing the denominator 2 Ď&#x192;4 by N i=1 (e â&#x2C6;&#x2019; Ď&#x192; robust to departures from normality.

14. Spatial Autocorrelation. Consider the regression model given in (9.1) with spatial autocorrelation deďŹ ned in (9.36). (a) Verify that the ďŹ rst-order conditions of maximization of the log-likelihood function given in (9.39) yield (9.41). (b) Show that for testing H0 ; Îť = 0, the score â&#x2C6;&#x201A;lnL/â&#x2C6;&#x201A;Îť evaluated under the null, i.e., at Îť = 0, is given by uâ&#x20AC;˛ W u/Ď&#x192;2 . (c) Show that the Information matrix with respect to Ď&#x192; 2 and Îť, evaluated under the null of Îť = 0, is given by â&#x17D;Ą â&#x17D;¤ n tr(W ) â&#x17D;˘ 2Ď&#x192; 4 â&#x17D;Ľ Ď&#x192;2 â&#x17D;˘ â&#x17D;Ľ â&#x17D;˘ â&#x17D;Ľ â&#x17D;Ł tr(W ) â&#x17D;Ś 2 â&#x20AC;˛ tr(W ) + tr(W W ) Ď&#x192;2 (d) Conclude from parts (b) and (c) that the Lagrange Multiplier for H0 ; Îť = 0 is given by LMÎť in (9.46). Hint: Use the fact that the diagonal elements of W are zero, hence tr(W ) = 0.

References Additional readings on GLS can be found in the econometric texts cited in the Preface. Anselin, L. (2001), â&#x20AC;&#x153;Spatial Econometrics,â&#x20AC;? Chapter 14 in B.H. Baltagi (ed.) A Companion to Theoretical Econometrics (Blackwell: Massachusetts). Anselin, L. (1988), Spatial Econometrics: Methods and Models (Kluwer: Dordrecht). Anselin, L. and A.K. Bera (1998), â&#x20AC;&#x153;Spatial Dependence in Linear Regression Models with an Introduction to Spatial Econometrics,â&#x20AC;? in A. Ullah and D.E.A. Giles (eds.) Handbook of Applied Economic Statistics (Marcel Dekker: New York). Balestra, P. (1970), â&#x20AC;&#x153;On the EďŹ&#x192;ciency of Ordinary Least Squares in Regression Models,â&#x20AC;? Journal of the American Statistical Association, 65: 1330-1337. Balestra, P. (1980), â&#x20AC;&#x153;A Note on the Exact Transformation Associated with First-Order Moving Average Process,â&#x20AC;? Journal of Econometrics, 14: 381-394. Baltagi, B.H. (1989), â&#x20AC;&#x153;Applications of a Necessary and SuďŹ&#x192;cient Condition for OLS to be BLUE,â&#x20AC;? Statistics and Probability Letters, 8: 457-461.


References

235

Baltagi, B.H. (1992), “Sampling Distributions and Efficiency Comparisons of OLS and GLS in the Presence of Both Serial Correlation and Heteroskedasticity,” Econometric Theory, Problem 92.2.3, 8: 304-305. Baltagi, B.H. and P.X. Wu (1997), “Estimation of Time Series Regressions with Autoregressive Disturbances and Missing Observations,” Econometric Theory, Problem 97.5.1, 13: 889. Baltagi, B.H. (1998), “Prediction in the Equicorrelated Regression Model,” Econometric Theory, Problem 98.3.3, 14: 382. Breusch, T.S. (1979), “Conflict Among Criteria for Testing Hypotheses: Extensions and Comments,” Econometrica, 47: 203-207. Breusch, T.S. and A.R. Pagan (1979), “A Simple Test for Heteroskedasticity and Random Coefficient Variation,” Econometrica, 47: 1287-1294. Buse, A. (1982), “The Likelihood Ratio, Wald, and Lagrange Multiplier Tests: An Expository Note,” The American Statistician, 36: 153-157. Dufour, J.M. (1986), “Bias of s2 in Linear Regressions with Dependent Errors,” The American Statistician, 40: 284-285. Fuller, W.A. and G.E. Battese (1974), “Estimation of Linear Models with Crossed-Error Structure,” Journal of Econometrics, 2: 67-78. Goldberger, A.S. (1962), “Best Linear Unbiased Prediction in the Generalized Linear Regression Model,” Journal of the American Statistical Association, 57: 369-375. Harvey, A.C. (1976), “Estimating Regression Models With Multiplicative Heteroskedasticity,” Econometrica, 44: 461-466. Im, E.I. and M.S. Snow (1993), “Sampling Distributions and Efficiency Comparisons of OLS and GLS in the Presence of Both Serial Correlation and Heteroskedasticity,” Econometric Theory, Solution 92.2.3, 9: 322-323. Kadiyala, K.R. (1968), “A Transformation Used to Circumvent the Problem of Autocorrelation,” Econometrica, 36: 93-96. Koenker, R. and G. Bassett, Jr. (1982), “Robust Tests for Heteroskedasticity Based on Regression Quantiles,” Econometrica, 50: 43-61. Kr¨ amer, W. and S. Berghoff (1991), “Consistency of s2 in the Linear Regression Model with Correlated Errors,” Empirical Economics, 16: 375-377. Kruskal, W. (1968), “When are Gauss-Markov and Least Squares Estimators Identical? A CoordinateFree Approach,” The Annals of Mathematical Statistics, 39: 70-75. Lempers, F.B. and T. Kloek (1973), “On a Simple Transformation for Second-Order Autocorrelated Disturbances in Regression Analysis,” Statistica Neerlandica, 27: 69-75. Magnus, J. (1978), “Maximum Likelihood Estimation of the GLS Model with Unknown Parameters in the Disturbance Covariance Matrix,” Journal of Econometrics, 7: 281-312. Milliken, G.A. and M. Albohali (1984), “On Necessary and Sufficient Conditions for Ordinary Least Squares Estimators to be Best Linear Unbiased Estimators,” The American Statistician, 38: 298299. Neudecker, H. (1977), “Bounds for the Bias of the Least Squares Estimator of σ 2 in Case of a First-Order Autoregressive Process (positive autocorrelation),” Econometrica, 45: 1257-1262.


236

Chapter 9: Generalized Least Squares

Neudecker, H. (1978), “Bounds for the Bias of the LS Estimator in the Case of a First-Order (positive) Autoregressive Process Where the Regression Contains a Constant Term,” Econometrica, 46: 12231226. Newey, W. and K. West (1987), “A Simple Positive Semi-Definite, Heteroskedasticity and Autocorrelation Consistent Covariance Matrix,” Econometrica, 55: 703-708. Oksanen, E.H. (1991), “A Simple Approach to Teaching Generalized Least Squares Theory,” The American Statistician, 45: 229-233. Ord, J.K. (1975), “ Estimation Methods for Models of Spatial Interaction,” Journal of the American Statistical Association, 70: 120-126. Phillips, P.C.B. and M.R. Wickens (1978), Exercises in Econometrics, Vol. 1 (Philip Allan/Ballinger: Oxford). Puntanen S. and G.P.H. Styan (1989), “The Equality of the Ordinary Least Squares Estimator and the Best Linear Unbiased Estimator,” (with discussion), The American Statistician, 43: 153-161. Sathe, S.T. and H.D. Vinod (1974), “Bounds on the Variance of Regression Coefficients Due to Heteroskedastic or Autoregressive Errors,” Econometrica, 42: 333-340. Schmidt, P. (1976), Econometrics (Marcell-Decker: New York). Termayne, A.R. (1985), “Prediction Error Variances Under Heteroskedasticity,” Econometric Theory, Problem 85.2.3, 1: 293-294. Theil, H. (1971), Principles of Econometrics (Wiley: New York). Thomas, J.J and K.F. Wallis (1971), “Seasonal Variation in Regression Analysis,” Journal of the Royal Statistical Society, Series A, 134: 67-72. White, H. (1980), “A Heteroskedasticity-Consistent Covariance Matrix Estimator and a Direct Test for Heteroskedasticity,” Econometrica, 48: 817-838. Zyskind, G. (1967), “On Canonical Forms, Non-Negative Covariance Matrices and Best and Simple Least Squares Linear Estimators in Linear Models,” The Annals of Mathematical Statistics, 38: 1092-1109.


CHAPTER 10

Seemingly Unrelated Regressions When asked “How did you get the idea for SUR?” Zellner responded: “On a rainy night in Seattle in about 1956 or 1957, I somehow got the idea of algebraically writing a multivariate regression model in single equation form. When I figured out how to do that, everything fell into place because then many univariate results could be carried over to apply to the multivariate system and the analysis of the multivariate system is much simplified notationally, algebraically and, conceptually.” Read the interview of Professor Arnold Zellner by Rossi (1989, p. 292).

10.1

Introduction

Consider two regression equations corresponding to two different firms yi = Xi β i + ui

i = 1, 2

(10.1)

where yi and ui are T × 1 and Xi is (T × Ki ) with ui ∼ (0, σ ii IT ). OLS is BLUE on each equation separately. Zellner’s (1962) idea is to combine these Seemingly Unrelated Regressions in one stacked model, i.e.,

    y1 X1 0 β1 u1 = + (10.2) y2 β2 u2 0 X2 which can be written as y = Xβ + u

(10.3)

where y ′ = (y1′ , y2′ ) and X and u are obtained similarly from (10.2). y and u are 2T × 1, X is 2T × (K1 + K2 ) and β is (K1 + K2 ) × 1. The stacked disturbances have a variance-covariance matrix

σ 12 IT σ 11 IT = Σ ⊗ IT (10.4) Ω= σ 21 IT σ 22 IT

√ where Σ = [σ ij ] for i, j = 1, 2; with ρ = σ 12 / σ 11 σ 22 measuring the extent of correlation between the two regression equations. The Kronecker product operator ⊗ is defined in the Appendix to Chapter 7. Some important applications of SUR models in economics include the estimation of a system of demand equations or a translog cost function along with its share equations, see Berndt (1991). Briefly, a system of demand equations explains household consumption of several commodities. The correlation among equations could be due to unobservable household specific attributes that influence the consumption of these commodities. Similarly, in estimating a cost equation along with the corresponding input share equations based on firm level data. The correlation among equations could be due to unobservable firm-specific effects that influence input choice and cost in production decisions.


238

Chapter 10: Seemingly Unrelated Regressions

Problem 1 asks the reader to verify that OLS on the system of two equations in (10.2) yields the same estimates as OLS on each equation in (10.1) taken separately. If ρ is large we expect gain in efficiency in performing GLS rather than OLS on (10.3). In this case ′ −1 −1 ′ −1  β GLS = (X Ω X) X Ω y

(10.5)

where Ω−1 = Σ−1 ⊗ IT . GLS will be BLUE for the system of two equations estimated jointly. Note that we only need to invert Σ to obtain Ω−1 . Σ is of dimension 2 × 2 whereas, Ω is of dimension 2T × 2T . In fact, if we denote by Σ−1 = [σ ij ], then GLS = β

σ 11 X1′ X1 σ 21 X2′ X1

σ 12 X1′ X2 σ 22 X2′ X2

−1

σ 11 X1′ y1 + σ 12 X1′ y2 σ 21 X2′ y1 + σ 22 X2′ y2

(10.6)

Zellner (1962) gave two sufficient conditions where it does not pay to perform GLS, i.e., GLS on this system of equations turns out to be OLS on each equation separately. These are the following: Case 1: Zero correlation among the disturbances of the i-th and j-th equations, i.e., σ ij = 0 for i = j. This means that Σ is diagonal which in turn implies that Σ−1 is diagonal with σ ii = 1/σ ii for i = 1, 2, and σ ij = 0 for i = j. Therefore, (10.6) reduces to 

 ′ ′ X )−1  β σ (X 0 X y /σ 11 1 1 11 1 1  β =  1,OLS (10.7) GLS = X2′ y2 /σ 22 0 σ 22 (X2′ X2 )−1 β 2,OLS

Case 2: Same regressors across all equations. This means that all the Xi ’s are the same, i.e.,

X1 = X2 = X ∗ . This rules out different number of regressors in each equation and all the Xi ’s must have the same dimension, i.e., K1 = K2 = K. Hence, X = I2 ⊗ X ∗ and (10.6) reduces to  β GLS

= [(I2 ⊗ X ∗′ )(Σ−1 ⊗ IT )(I2 ⊗ X ∗ )]−1 [(I2 ⊗ X ∗′ )(Σ−1 ⊗ IT )y] ∗′

∗ −1

= [Σ ⊗ (X X )

][(Σ

−1

∗′

∗′

∗ −1

⊗ X )y] = [I2 ⊗ (X X )

(10.8)

 X ]y = β OLS ∗′

These results generalize to the case of M regression equations, but for simplicity of exposition we considered the case of two equations only. A necessary and sufficient condition for SUR(GLS) to be equivalent to OLS, was derived by Dwivedi and Srivastava (1978). An alternative derivation based on the Milliken and Albohali (1984) necessary and sufficient condition for OLS to be equivalent to GLS, is presented here, see Baltagi (1988). In Chapter 9, we saw that GLS is equivalent to OLS, for every y, if and only if X ′ Ω−1 P¯X = 0

(10.9)

In this case, X = diag[Xi ], Ω−1 = Σ−1 ⊗ IT , and P¯X = diag[P¯Xi ]. Hence, the typical element of (10.9), see problem 1, is σ ij Xi′ P¯Xj = 0

(10.10)

This is automatically satisfied for i = j. For i = j, this holds if σ ij = 0 or Xi′ P¯Xj = 0. Note that σ ij = 0 is the first sufficient condition provided by Zellner (1962). The latter condition Xi′ P¯Xj = 0 implies that the set of regressors in the i-th equation are a perfect linear combination


10.2

Feasible GLS Estimation

239

of those in the j-th equation. Since Xj′ P¯Xi = 0 has to hold also, Xj has to be a perfect linear combination of the regressors in the i-th equation. Xi and Xj span the same space. Both Xi and Xj have full column rank for OLS to be feasible, hence they have to be of the same dimension for Xi′ P¯Xj = Xj′ P¯Xi = 0. In this case, Xi′ = CXj′ , where C is a nonsingular matrix, i.e., the regressors in the i-th equation are a perfect linear combination of those in the j-th equation. This includes the second sufficient condition derived by Zellner (1962). In practice, different economic behavioral equations contain different number of right hand side variables. In this case, one rearranges the SUR into blocks where each block has the same number of right hand side variables. For two equations (i and j) belonging to two different blocks (i = j), (10.10) is satisfied if the corresponding σ ij is zero, i.e., Σ has to be block diagonal. However, in this case, GLS performed on the whole system is equivalent to GLS performed on each block taken separately. Hence, (10.10) is satisfied for SUR if it is satisfied for each block taken separately. Revankar (1974) considered the case where X2 is a subset of X1 . In this case, there is no gain in   using SUR for estimating β 2 . In fact, problem 2 asks the reader to verify that β 2,SU R = β 2,OLS .   However, this is not the case for β 1 . It is easy to show that β 1,SU R = β 1,OLS − Ae2,OLS , where A is a matrix defined in problem 2, and e2,OLS are the OLS residuals for the second equation. Telser (1964) suggested an iterative least squares procedure for SUR equations. For the two equations model given in (10.1), this estimation method involves the following: 1. Compute the OLS residuals e1 and e2 from both equations. 2. Include e1 as an extra regressor in the second equation and e2 as an extra regressor in the first equation. Compute the new least squares residuals and iterate this step until convergence of the estimated coefficients. The resulting estimator has the same asymptotic distribution as Zellner’s (1962) SUR estimator. Conniffe (1982) suggests stopping at the second step because in small samples this provides most of the improvement in precision. In fact, Conniffe (1982) argues that it may be unnecessary and even disadvantageous to calculate Zellner’s estimator proper. Extensions to multiple equations is simple. Step 1 is the same where one computes least squares residuals of every equation. Step 2 adds the residuals of all other equations in the equation of interest. OLS is run and the new residuals are computed. One can stop at this second step or iterate until convergence.

10.2

Feasible GLS Estimation

In practice, Σ is not known and has to be estimated. Zellner (1962) recommended the following feasible GLS estimation procedure:  sii = Tt=1 e2it /(T − Ki ) for i = 1, 2 (10.11)

and

sij =

T

t=1 eit ejt /(T

− Ki )1/2 (T − Kj )1/2

for i, j = 1, 2

and i = j

(10.12)

where eit denotes OLS residuals of the i-th equation. sii is the s2 of the regression for the i-th equation. This is unbiased for σ ii . However, sij for i = j is not unbiased for σ ij . In fact, the unbiased estimate is  for i, j = 1, 2 (10.13) sij = Tt=1 eit ejt /[T − Ki − Kj + tr (B)]


240

Chapter 10: Seemingly Unrelated Regressions

where B = Xi (Xi′ Xi )−1 Xi′ Xj (Xj′ Xj )−1 Xj′ = PXi PXj , see problem 4. Using this last estimator may lead to a variance-covariance matrix that is not positive definite. For consistency, however, all we need is a division by T , however this leaves us with a biased estimator:  sij = Tt=1 eit ejt /T for i, j = 1, 2 (10.14)

Using this consistent estimator of Σ will result in feasible GLS estimates that are asymptotically efficient. In fact, if one iterates this procedure, i.e., compute feasible GLS residuals and second round estimates of Σ using these GLS residuals in (10.14), and continue iterating, until convergence, this will lead to maximum likelihood estimates of the regression coefficients, see Oberhofer and Kmenta (1974). Relative Efficiency of OLS in the Case of Simple Regressions To illustrate the gain in efficiency of Zellner’s SUR compared to performing OLS on each equation separately, Kmenta(1986, pp.641-643) considers the following two simple regression equations: Y1t = β 11 + β 12 X1t + u1t Y2t = β 21 + β 22 X2t + u2t

(10.15) for

t = 1, 2, . . . T ;

and proves that 2 2 2   var(β 12,GLS )/var(β 12,OLS ) = (1 − ρ )/[1 − ρ r ]

(10.16)

where ρ is the correlation coefficient between u1 and u2 , and r is the sample correlation coefficient between X1 and X2 . Problem 5 asks the reader to verify (10.16). In fact, the same relative   efficiency ratio holds for β 22 , i.e., var(β 22,GLS )/var(β 22,OLS ) is given by that in (10.16). This confirms the two results obtained above, namely, that as ρ increases this relative efficiency ratio decreases and OLS is less efficient than GLS. Also, as r increases this relative efficiency ratio increases and there is less gain in performing GLS rather than OLS. For ρ = 0 or r = 1, the efficiency ratio is 1, and OLS is equivalent to GLS. However, if ρ is large, say 0.9 and r is small, say 0.1 then (10.16) gives a relative efficiency of 0.11. For a tabulation of (10.16) for various values of ρ2 and r 2 , see Table 12-1 of Kmenta (1986, p. 642). Relative Efficiency of OLS in the Case of Multiple Regressions With more regressors in each equation, the relative efficiency story has to be modified, as indicated by Binkley and Nelson (1988). In the two equation model considered in (10.2) with K1 regressors X1 in the first equation and K2 regressors X2 in the second equation

11 ′ −1 σ 12 X1′ X2  GLS ) = (X ′ Ω−1 X)−1 = σ X1 X1 = [Aij ] (10.17) var(β σ 22 X2′ X2 σ 21 X2′ X1

 If we focus on the regression estimates of the first equation, we get var(β 1,GLS ) = A11 = −1 −1 21 ′ 22 ′ 12 ′ 11 ′ [σ X1 X1 − σ X1 X2 (σ X2 X2 ) σ X2 X1 ] see problem 6. Using the fact that

1/σ 11 −ρ2 /σ 12 Σ−1 = [1/(1 − ρ2 )] −ρ2 /σ 21 1/σ 22


10.2

Feasible GLS Estimation

241

where ρ2 = σ 212 /σ 11 σ 22 , one gets 2 ′ 2 ′ −1  var(β 1,GLS ) = [σ 11 (1 − ρ )]{X1 X1 − ρ (X1 PX2 X1 )}

(10.18)

′ 2 2 ′ −1  var(β 1,GLS ) = σ 11 {X1 X1 + [ρ /(1 − ρ )]E E}

(10.19)

Add and subtract ρ2 X1′ X1 from the expression to be inverted, one gets

where E = P¯X2 X1 is the matrix whose columns are the OLS residuals of each variable in X1 regressed on X2 . If E = 0, there is no gain in SUR over OLS for the estimation of β 1 . X1 = X2 or X1 is a subset of X2 are two such cases. One can easily verify that (10.19) is the variance-covariance matrix of an OLS regression with regressor matrix

X1 W = θE where θ2 = ρ2 /(1 − ρ2 ). Now let us focus on the efficiency of the estimated coefficient of the q-th variable, Xq in X1 . Recall, from Chapter 4, that for the regression of y on X1 3 2  q,OLS ) = σ 11 / T x2 (1 − R2 ) (10.20) var(β q t=1 tq

where the denominator is the residual sum of squares of Xq on the other (K1 − 1) regressors in X1 and Rq2 is the corresponding R2 of that regression. Similarly, from (10.19), 2 3 T ∗2 2 2 2 T  (10.21) e x + θ var(β ) = σ / 11 q,SU R t=1 tq (1 − Rq ) t=1 tq

Xq on the other (K1 −1) regressors θeq  in W , and Rq∗2 is the corresponding R2 of that regression. Add and subtract Tt=1 x2tq (1 − Rq2 ) to the denominator of (10.21), we get where the denominator is the residual sum of squares of

 var(β q,SU R ) = 2T

2 t=1 xtq (1

Rq2 ) +

T

σ 11

2 2 t=1 xtq (Rq

− Rq∗2 ) + θ2

3 (10.22) 2 (1 − R∗2 ) e q t=1 tq

T

 This variance differs from var(β q,OLS ) in (10.20) by the two extra terms in the denominator. If 2 ′ ρ = 0, then θ = 0, so that W = [X1′ , 0] and Rq2 = Rq∗2 . In this case, (10.22) reduces to (10.20). If Xq also appears in the second equation, or in general is spanned by the variables in X2 , then  etq = 0, Tt=1 e2tq = 0 and from (10.22) there is gain in efficiency only if Rq2 ≥ Rq∗2 . Rq2 is a measure of multicollinearity of Xq with the other (K1 −1) regressors in the first equation, i.e., X1 . If this is high, then it is more likely for Rq2 ≥ Rq∗2 . Therefore, the higher the multicollinearity within X1 , the greater the potential for a decrease in variance of OLS by SUR. Note that Rq2 = Rq∗2 when θE = 0. This is true if θ = 0, or E = 0. The latter occurs when X1 is spanned by the sub-space of X2 . Problem 7 asks the reader to verify that Rq2 = Rq∗2 when X1 is orthogonal to X2 . Therefore, with more regressors in each equation, one has to consider the correlation between the X’s within each equation as well as that across equations. Even when the X’s across equations are highly correlated, there may still be gains from joint estimation using SUR when there is high mulicollinearity within each equation.


242

Chapter 10: Seemingly Unrelated Regressions

10.3

Testing Diagonality of the Variance-Covariance Matrix

Since the diagonality of ÎŁ is at the heart of using SUR estimation methods, it is important to look at tests for H0 : ÎŁ is diagonal. Breusch and Pagan (1980) derived a simple and easy to use Lagrange multiplier statistic for testing H0 . This is based upon the sample correlation coeďŹ&#x192;cients of the OLS residuals: LM = T

M iâ&#x2C6;&#x2019;1 i=2

2 j=1 rij

(10.23)

where M denotes the number of equations and rij = sij /( sii sjj )1/2 . The sij â&#x20AC;&#x2122;s are computed from OLS residuals as in (10.14). Under the null hypothesis, ÎťLM has an asymptotic Ď&#x2021;2M (M â&#x2C6;&#x2019;1)/2 distribution. Note that the sij â&#x20AC;&#x2122;s are needed for feasible GLS estimation. Therefore, it is easy to compute the rij â&#x20AC;&#x2122;s and ÎťLM by summing the squares of half the number of off-diagonal elements of R = [rij ] and multiplying the sum by T . For example, for the two equations case, 2 which is asymptotically distributed as Ď&#x2021;2 under H . For the three equations case, ÎťLM = T r21 0 1 2 + r 2 + r 2 ) which is asymptotically distributed as Ď&#x2021;2 under H . ÎťLM = T (r21 0 3 31 32 Alternatively, the Likelihood Ratio test can also be used to test for diagonality of ÎŁ. This is based on the determinants of the variance covariance matrices estimated by MLE for the restricted and unrestricted models:   M  log s â&#x2C6;&#x2019; log| ÎŁ| (10.24) ÎťLR = T ii i=1

where sii is the restricted MLE of Ď&#x192; ii obtained from the OLS residuals as in (10.14). The matrix  denotes the unrestricted MLE of ÎŁ. This may be adequately approximated with an estimator ÎŁ  based on the feasible GLS estimator β F GLS , see Judge et al. (1982). Under H0 , ÎťLR has an 2 asymptotic Ď&#x2021;M (M â&#x2C6;&#x2019;1)/2 distribution.

10.4

Seemingly Unrelated Regressions with Unequal Observations

Srivastava and Dwivedi (1979) surveyed the developments in the SUR model and described the extensions of this model to the serially correlated case, the nonlinear case, the misspeciďŹ ed case, and that with unequal number of observations. Srivastava and Giles (1988) dedicated a monograph to SUR models, and surveyed the ďŹ nite sample as well as asymptotic results. More recently, Fiebig (2001) gives a concise and up to date account of research in this area. In this section, we consider one extension to focus upon. This is the case of SUR with unequal number of observations considered by Schmidt (1977), Baltagi, Garvin and Kerman (1989) and Hwang (1990). Let the ďŹ rst ďŹ rm have T observations common with the second ďŹ rm, but allow the latter to have N extra observations. In this case, (10.2) will have y1 of dimension T Ă&#x2014; 1 whereas y2 will be of dimension (T + N ) Ă&#x2014; 1. In fact, y2â&#x20AC;˛ = (y2â&#x2C6;&#x2014;â&#x20AC;˛ , y2oâ&#x20AC;˛ ) and X2â&#x20AC;˛ = (X2â&#x2C6;&#x2014;â&#x20AC;˛ , X2oâ&#x20AC;˛ ) with â&#x2C6;&#x2014; denoting the T common observations for the second ďŹ rm, and o denoting the extra N observations for the second ďŹ rm. The disturbances will now have a variance-covariance matrix â&#x17D;¤ â&#x17D;Ą 0 Ď&#x192; 11 IT Ď&#x192; 12 IT â&#x17D;Ś (10.25) Ί = â&#x17D;Ł Ď&#x192; 12 IT Ď&#x192; 22 IT 0 0 0 Ď&#x192; 22 IN


10.4

Seemingly Unrelated Regressions with Unequal Observations

243

GLS on (10.2) will give  β GLS

=

σ 11 X1′ X1 σ 12 X1′ X2∗ 12 ∗′ 22 ∗′ σ X2 X1 σ X2 X2∗ + (X2o′ X2o )/σ 22 )

σ 11 X1′ y1 + σ 12 X1′ y2∗ σ 12 X2∗′ y1 + σ 22 X2∗′ y2∗

+ (X2o′ y2o )/σ 22 )

−1

(10.26)

where Σ−1 = [σ ij ] for i, j = 1, 2. If we run OLS on each equation (T for the first equation, and T + N for the second equation) and denote the residuals for the two equations by e1 and e2 , o′ respectively, then we can partition the latter residuals into e′2 = (e∗′ 2 , e2 ). In order to estimate Ω, Schmidt (1977) considers the following procedures: (1) Ignore the extra N observations in estimating Ω. In this case 12 = s12 = e′1 e∗2 /T σ 11 = s11 = e′1 e1 /T ; σ

∗ and σ 22 = s∗22 = e∗′ 2 e2 /T

(10.27)

22 = s22 = (2) Use T + N observations to estimate σ 22 . In other words, use s11 , s12 and σ e′2 e2 /(T + N ). This procedure is attributed to Wilks (1932) and has the disadvantage of giving estimates of Ω that are not positive definite.  is positive definite. Srivastava (3) Use s11 and s22 , but modify the estimate of σ 12 such that Ω 1/2 ∗ and Zaatar (1973) suggest σ 12 = s12 (s22 /s22 ) .

(4) Use all (T + N ) observations in estimating Ω. Hocking and Smith (1968) suggest using o σ 11 = s11 − (N/N + T )(s12 /s∗22 )2 (s∗22 − so22 ) where so22 = eo′ 12 = s12 (s22 /s∗22 ) and 2 e2 /N ; σ σ 22 = s22 . (5) Use a maximum likelihood procedure.

 F GLS based on any of these estimators will be asympAll estimators of Ω are consistent, and β totically efficient. Schmidt considers their small sample properties by means of Monte Carlo experiments. Using the set up of Kmenta and Gilbert (1968) he finds for T = 10, 20, 50 and N = 5, 10, 20 and various correlation of the X’s and the disturbances across equations the following disconcerting result: “..it is certainly remarkable that procedures that essentially ignore the extra observations in estimating Σ (e.g., Procedure 1) do not generally do badly relative to procedures that use the extra observations fully (e.g., Procedure 4 or MLE). Except when the disturbances are highly correlated across equations, we may as well just forget about the extra observations in estimating Σ. This is not an intuitively reasonable procedure.” Hwang (1990) re-parametrizes these estimators in terms of the elements of Σ−1 rather than Σ. After all, it is Σ−1 rather than Σ that appears in the GLS estimator of β. This re-parametrization shows that the estimators of Σ−1 no longer have the ordering in terms of their use of the extra observations as that reported by Schmidt (1977). However, regardless of the parametrization chosen, it is important to point out that all the observations are used in the estimation of β whether at the first stage for obtaining the least squares residuals, or in the final stage in computing GLS. Baltagi et al. (1989) show using Monte Carlo experiments that better estimates of Σ or its inverse Σ−1 in Mean Square Error sense, do not necessarily lead to better GLS estimates of β.


244

Chapter 10: Seemingly Unrelated Regressions

10.5

Empirical Example

Baltagi and Griffin (1983) considered the following gasoline demand equation:

log

Y PM G Car Gas = α + β 1 log + β 2 log + β 3 log +u Car N PGDP N

where Gas/Car is motor gasoline consumption per auto, Y/N is real per capita income, PM G /PGDP is real motor gasoline price and Car/N denotes the stock of cars per capita. This data consists of annual observations across 18 OECD counties, covering the period 19601978. It is provided as GASOLINE.DAT on the Springer web site. We consider the first two countries: Austria and Belgium. OLS on this data yields Austria

log

Gas Car

=

Y PM G Car − 0.520 log 3.727 + 0.761 log − 0.793 log N PGDP N (0.113) (0.211) (0.373) (0.150)

Belgium

log

Gas Car

=

Y PM G Car − 0.673 log 3.042 + 0.845 log − 0.042 log N PGDP N (0.453) (0.170) (0.158) (0.093)

where the standard errors are shown in parentheses. Based on these OLS residuals, the estimate of Σ is given by = Σ

0.0012128 0.00023625 0.00092367

 i.e., after one iteration, are The Seemingly Unrelated Regression estimates based on this Σ, given by Austria

log

Gas Car

=

Y PM G Car − 0.496 log 3.713 + 0.721 log − 0.754 log N PGDP N (0.209) (0.372) (0.146) (0.111)

Belgium

log

Gas Car

=

Y PM G Car − 0.686 log 2.843 + 0.835 log − 0.131 log N PGDP N (0.445) (0.170) (0.154) (0.093)

2 = 0.947 which is The Breusch-Pagan (1980) Lagrange multiplier test for diagonality of Σ is T r21 2 distributed as χ1 under the null hypothesis. The Likelihood Ratio test for the diagonality of Σ, given in (10.23), yields a value of 1.778 which is also distributed as χ21 under the null hypothesis. Both test statistics do not reject H0 . These SUR results were run using SHAZAM and could be iterated further. Note the reduction in the standard errors of the estimated regression coefficients is minor as we compare the OLS and SUR estimates. Suppose that we only have the first 15 observations (1960-1974) on Austria and all 19 observations (1960-1978) on Belgium. We now apply the four feasible GLS procedures described by Schmidt (1977). The first procedure which ignores the extra 4 observations in estimating Σ yields s11 = 0.00086791, s12 = 0.00026357 and s∗22 = 0.00109947 as described in (10.27). The


Problems

245

resulting SUR estimates are given by Y PM G Car Gas = − 0.487 log Austria log 4.484 + 0.817 log − 0.580 log Car N PGDP N (0.438) (0.168) (0.176) (0.098) Belgium

log

Gas Car

=

Y PM G Car − 0.686 log 2.936 + 0.848 log − 0.095 log N PGDP N (0.164) (0.436) (0.151) (0.090)

The second procedure, due to Wilks (1932) uses the same s11 and s12 in procedure 1, but σ 22 = s22 = e′2 e2 /19 = 0.00092367. The resulting SUR estimates are given by Austria

log

Gas Car

=

Y PM G Car − 0.476 log 4.521 + 0.806 log − 0.554 log N PGDP N (0.437) (0.167) (0.174) (0.098)

Belgium

log

Gas Car

=

Y PM G Car − 0.685 log 2.937 + 0.848 log − 0.094 log N PGDP N (0.399) (0.150) (0.138) (0.082)

The third procedure based on Srivastava and Zaatar (1973) use the same s11 and s22 as procedure 2, but modify σ 12 = s12 (s22 /s∗22 )1/2 = 0.00024158. The resulting SUR estimates are given by Y PM G Car Gas = − 0.481 log Austria log 4.503 + 0.812 log − 0.567 log Car N PGDP N (0.168) (0.438) (0.176) (0.098) Belgium

log

Gas Car

=

Y PM G Car − 0.684 log 2.946 + 0.847 log − 0.090 log N PGDP N (0.400) (0.151) (0.139) (0.082)

12 = The fourth procedure due to Hocking and Smith (1968) yields σ 11 = 0.00085780, σ 0.0022143 and σ 22 = s22 = 0.00092367. The resulting SUR estimates are given by Y PM G Car Gas = − 0.487 log Austria log 4.485 + 0.817 log − 0.579 log Car N PGDP N (0.168) (0.437) (0.176) (0.098) Belgium

log

Gas Car

=

Y PM G Car − 0.684 log 2.952 + 0.847 log − 0.086 log N PGDP N (0.151) (0.400) (0.139) (0.082)

In this case, there is not much difference among these four alternative estimates.

Problems 1.

(a) Show that OLS on a system of two Zellner’s SUR equations given in (10.2) is the same as OLS on each equation taken separately. What about the estimated variance-covariance matrix of the coefficients? Will they be the same? (b) In the General Linear Model, we found a necessary and sufficient condition for OLS to be equivalent to GLS is that X ′ Ω−1 P¯X = 0 for every y where P¯X = I − PX . Show that a necessary and sufficient condition for Zellner’s GLS to be equivalent to OLS is that σ ij Xi′ P¯Xj = 0 for i = j as described in (10.10). This is based on Baltagi (1988). (c) Show that the two sufficient conditions given by Zellner for SUR to be equivalent to OLS both satisfy the necessary and sufficient condition given in part (b).


246

Chapter 10: Seemingly Unrelated Regressions (d) Show that if Xi = Xj C â&#x20AC;˛ where C is an arbitrary nonsingular matrix, then the necessary and suďŹ&#x192;cient condition given in part (b) is satisďŹ ed.

2. For the two SUR equations given in (10.2). Let X1 = (X2 , Xe ), i.e., X2 is a subset of X1 . Prove that   (a) β 2,SUR = β 2,OLS .

  (b) β 12 (X1â&#x20AC;˛ X1 )â&#x2C6;&#x2019;1 X1â&#x20AC;˛ / s22 . e2,OLS are the OLS residuals 1,SUR = β 1,OLS â&#x2C6;&#x2019; Ae2,OLS , where A = s from the second equation, and the sij â&#x20AC;&#x2122;s are deďŹ ned in (10.14).

3. For the two SUR equations given in (10.2). Let X1 and X2 be orthogonal, i.e., X1â&#x20AC;˛ X2 = 0. Show that knowing the true ÎŁ we get 12 11 â&#x20AC;˛ â&#x2C6;&#x2019;1 â&#x20AC;˛ 21 22 â&#x20AC;˛ â&#x2C6;&#x2019;1     X1 y2 and β (a) β 1,GLS = β 1,OLS + (Ď&#x192; /Ď&#x192; )(X1 X1 ) 2,GLS = β 2,OLS + (Ď&#x192; /Ď&#x192; )(X2 X2 ) â&#x20AC;˛ X2 y1 .

(b) What are the variances of these estimates?

 (c) If X1 and X2 are single regressors, what are the relative eďŹ&#x192;ciencies of β i,OLS with respect  to β i,GLS for i = 1, 2?

4. Verify that sij , given in (10.13), is unbiased for Ď&#x192; ij . Note that for computational purposes tr(B) = tr(PXi PXj ).

5. Relative Efficiency of OLS in the Case of Simple Regressions. This is based on Kmenta (1986, pp. 641-643). For the system of two equations given in (10.15), show that T   (a) var(β 12,OLS ) = Ď&#x192; 11 /mx1 x1 and var(β 22,OLS ) = Ď&#x192; 22 /mx2 x2 where mxi xj = t=1 (Xit â&#x2C6;&#x2019; ÂŻ i )(Xjt â&#x2C6;&#x2019; X ÂŻ j ) for i, j = 1, 2. X !

â&#x2C6;&#x2019;1  β Ď&#x192; 22 mx1 x1 â&#x2C6;&#x2019;Ď&#x192; 12 mx1 x2 12,GLS . = (Ď&#x192; 11 Ď&#x192; 22 â&#x2C6;&#x2019; Ď&#x192; 212 ) (b) var  â&#x2C6;&#x2019;Ď&#x192; 12 mx1 x2 Ď&#x192; 11 mx2 x2 β 22,GLS   2 2 2  Deduce that var β 12,GLS = (Ď&#x192; 11 Ď&#x192; 22 â&#x2C6;&#x2019; Ď&#x192; 12 )Ď&#x192; 11 mx2 x2 /[Ď&#x192; 11 Ď&#x192; 22 mx2 x2 mx1 x1 â&#x2C6;&#x2019; Ď&#x192; 12 mx1 x2 ] and   2 2 2  var β 22,GLS = (Ď&#x192; 11 Ď&#x192; 22 â&#x2C6;&#x2019; Ď&#x192; 12 )Ď&#x192; 22 mx1 x1 /[Ď&#x192; 11 Ď&#x192; 22 mx1 x1 mx2 x2 â&#x2C6;&#x2019; Ď&#x192; 12 mx1 x2 ].

(c) Using Ď = Ď&#x192; 12 /(Ď&#x192; 11 Ď&#x192; 22 )1/2 and r = mx1 x2 /(mx1 x1 mx2 x2 )1/2 and the results in parts (a) and 2 2 2   (b), show that (10.16) holds, i.e., var(β 12,GLS )/var(β 12,OLS ) = (1 â&#x2C6;&#x2019; Ď )/[1 â&#x2C6;&#x2019; Ď r ].

(d) Differentiate (10.16) with respect to θ = Ď 2 and show that (10.16) is a non-increasing function of θ. Similarly, differentiate (10.16) with respect to Îť = r2 and show that (10.16) is a nondecreasing function of Îť. Finally, compute this eďŹ&#x192;ciency measure (10.16) for various values of Ď 2 and r2 between 0 and 1 at 0.1 intervals, see Kmentaâ&#x20AC;&#x2122;s (1986) Table 12-1, p.642. 6. Relative Efficiency of OLS in the Case of Multiple Regressions. This is based on Binkley and  Nelson (1988). Using partitioned inverse formulas, verify that var(β 1,GLS ) = A11 given below (10.17). Deduce (10.18) and (10.19).

7. Consider the multiple regression case with orthogonal regressors across the two equations, i.e., X1â&#x20AC;˛ X2 = 0. Verify that Rq2 = Rqâ&#x2C6;&#x2014;2 , where Rq2 and Rqâ&#x2C6;&#x2014;2 are deďŹ ned below (10.20) and (10.21), respectively.

8.

(a) SUR With Unequal Number of Observations. This is based on Schmidt (1977). Derive the GLS estimator for SUR with unequal number of observations given in (10.26). (b) Show that if Ď&#x192; 12 = 0, SUR with unequal number of observations reduces to OLS on each equation separately.


Problems

247

9. Grunfeld (1958) considered the following investment equation: Iit = α + β 1 Fit + β 2 Cit + uit where Iit denotes real gross investment for firm i in year t, Fit is the real value of the firm (shares outstanding) and Cit is the real value of the capital stock. This data set consists of 10 large U.S. manufacturing firms over 20 years, 1935-1954, and are given in Boot and de Witt (1960). It is provided as GRUNFELD.DAT on the Springer web site. Consider the first three firms: G.M., U.S. Steel and General Electric. (a) Run OLS of I on a constant, F and C for each of the 3 firms separately. Plot the residuals against time. Print the variance-covariance of the estimates. (b) Test for serial correlation in each regression. (c) Run Seemingly Unrelated Regressions (SUR) for the first two firms. Compare with OLS. (d) Run SUR for the three assigned firms. Compare these results with those in part (c). (e) Test for the diagonality of Σ across these three equations. (f) Test for the equality of all coefficients across the 3 firms. 10. (Continue problem 9). Consider the first two firms again and focus on the coefficient of F . Refer to the Binkley and Nelson (1988) article in The American Statistician, and compute Rq2 , Rq∗2 , Σe2tq and Σx2tq . (a) What would be equations (10.20) and (10.21) for your data set? (b) Substitute estimates of σ 11 and θ2 and verify that the results are the same as those obtained in problems 9(a) and 9(c). (c) Compare the results from equations (10.20) and (10.21) in part (a). What do you conclude? 11. (Continue problem 9). Consider the first two firms once more. Now you only have the first 15 observations on the first firm and all 20 observations on the second firm. Apply Schmidt’s (1977) feasible GLS estimators and compare the resulting estimates. 12. For the Baltagi and Griffin (1983) Gasoline Data considered in section 10.5, the model is log

Y PMG Car Gas = α + β 1 log + β 2 log +u + β 3 log Car N PGDP N

where Gas/Car is motor gasoline consumption per auto, Y/N is real per capita income, PMG /PGDP is real motor gasoline price and Car/N denotes the stock of cars per capita. (a) Run Seemingly Unrelated Regressions (SUR) for the first two countries. Compare with OLS. (b) Run SUR for the first three countries. Comment on the results and compare with those of part (a). (Are there gains in efficiency?) (c) Test for Diagonality of Σ across the three equations using the Breusch and Pagan (1980) LM test and the Likelihood Ratio test. (d) Test for the equality of all coefficients across the 3 countries. (e) Consider the first 2 countries once more. Now you only have the first 15 observations on the first country and all 19 observations on the second country. Apply Schmidt’s (1977) feasible GLS estimators, and compare the results.


248

Chapter 10: Seemingly Unrelated Regressions

13. Trace Minimization of Singular Systems with Cross-Equation Restrictions. This is based on Baltagi (1993). Berndt and Savin (1975) demonstrated that when certain cross-equation restrictions are imposed, restricted least squares estimation of a singular set of SUR equations will not be invariant to which equation is deleted. Consider the following set of three equations with the same regressors: yi = αi ιT + β i X + ǫi

i = 1, 2, 3.

where yi = (yi1 , yi2 , . . . , yiT )′ , X = (x1 , x2 , . . . , xT )′ , and ǫi for (i = 1, 2, 3) are T × 1 vectors and ιT is a vector of  ones of dimension T . αi and β i are scalars, and these equations satisfy the 3 adding up restriction i=1 yit = 1 for every t = 1, 2, . . . , T . Additionally, we have a cross-equation restriction: β 1 = β 2 . T T ¯)2 ¯)yit / t=1 (xt − x (a) Denote the unrestricted OLS estimates of β i by bi where bi = t=1 (xt − x T for i = 1, 2, 3, and x¯ = t=1 xt /T . Show that these unrestricted bi ’s satisfy the adding up restriction β 1 + β 2 + β 3 = 0 on the true parameters automatically. (b) Show that if one drops the first equation for i = 1 and estimate the remaining system by  = 0.4b1 + 0.6b2 . trace minimization subject to β 1 = β 2 , one gets β 1

(c) Now drop the second equation for i = 2, and show that estimating the remaining system by  = 0.6b1 + 0.4b2 . trace minimization subject to β 1 = β 2 , gives β 1

(d) Finally, drop the third equation for i = 3, and show that estimating the remaining system  = 0.5b1 + 0.5b2 . by trace minimization subject to β 1 = β 2 gives β 1

 is not invariant to the deleted equation. Also, this Note that this also means the variance of β 1 non-invariancy affects Zellner’s SUR estimation if the restricted least squares residuals are used rather than the unrestricted least squares residuals in estimating the variance covariance matrix of the disturbances. Hint: See the solution by Im (1994). 14. For the Natural Gas data considered in Chapter 4, problem 16. The model is logConsit

= β 0 + β 1 logP git + β 2 logP oit + β 3 logP eit + β 4 logHDDit +β 5 logP Iit + uit

where i = 1, 2, . . . , 6 states and t = 1, 2, . . . , 23 years. (a) Run Seemingly Unrelated Regressions (SUR) for the first two states. Compare with OLS. (b) Run SUR for all six states. Comment on the results and compare with those of part (a). (Are there gains in efficiency?) (c) Test for Diagonality of Σ across the six states using the Breusch and Pagan (1980) LM test and the Likelihood Ratio test. (d) Test for the equality of all coefficients across the six states. 15. Equivalence of LR Test and Hausman Test. This is based on Qian (1998). Suppose that we have the following two equations: ygt = αg + ugt

g = 1, 2,

t = 1, 2, . . . T

where (u1t , u2t ) is normally distributed with mean zero and variance Ω = Σ ⊗ IT where Σ = [σ gs ] for g, s = 1, 2. This is a simple example of the same regressors across two equations. (a) Show that the OLS estimator of αg is the same as the GLS estimator of αg and both are T equal to y¯g = t=1 ygt /T for g = 1, 2.


References

249

(b) Derive the maximum likelihood estimators of αg and σ gs for g, s, = 1, 2. Compute the loglikelihood function evaluated at these unrestricted estimates. (c) Compute the maximum likelihood estimators of αg and σ gs for g, s = 1, 2 under the null hypothesis H0 ; σ 11 = σ 22 . (d) Using parts (b) and (c) compute the LR test for H0 ; σ 11 = σ 22 . (e) Show that the LR test for H0 derived in part (c) is asymptotically equivalent to the Hausman test based on the difference in estimators obtained in parts (b) and (c). Hausman’s test is studied in Chapter 12. 16. Estimation of a Triangular, Seemingly Unrelated Regression System by OLS. This is based on Sentana (1997). Consider a system of three SUR equations in which the explanatory variables for the first equation are a subset of the explanatory variables for the second equation, which are in turn a subset of the explanatory variables for the third equation. (a) Show that SUR applied to the first two equations is the same (for those equations) as SUR applied to all three equations. Hint: See Schmidt (1978). (b) Using part (a) show that SUR for the first equation is equivalent to OLS. (c) Using parts (a) and (b) show that SUR for the second equation is equivalent to OLS on the second equation with one additional regressor. The extra regressor is the OLS residuals from the first equation. Hint: Use Telser’s (1964) results. (d) Using parts (a), (b) and (c) show that SUR for the third equation is equivalent to OLS on the third equation with the residuals from the regressions in parts (b) and (c) as extra regressors.

References This chapter is based on Zellner(1962), Kmenta(1986), Baltagi (1988), Binkley and Nelson (1988), Schmidt (1977) and Judge et al. (1982). References cited are: Baltagi, B.H. (1988), “The Efficiency of OLS in a Seemingly Unrelated Regressions Model,” Econometric Theory, Problem 88.3.4, 4: 536-537. Baltagi, B.H. (1993), “Trace Minimization of Singular Systems With Cross-Equation Restrictions,” Econometric Theory, Problem 93.2.4, 9: 314-315. Baltagi, B., S. Garvin and S. Kerman (1989), “Further Monte Carlo Evidence on Seemingly Unrelated Regressions with Unequal Number of Observations,” Annales D ′ Economie et de Statistique, 14: 103-115. Baltagi, B.H. and J.M. Griffin (1983), “Gasoline Demand in the OECD: An Application of Pooling and Testing Procedures,” European Economic Review, 22: 117-137. Berndt, E.R. (1991), The Practice of Econometrics: Classic and Contemporary (Addison- Wesley: Reading, MA). Berndt, E.R. and N.E. Savin (1975), “Estimation and Hypothesis Testing in Singular Equation Systems With Autoregressive Disturbances,” Econometrica, 43: 937-957. Binkley, J.K. and C.H. Nelson (1988), “A Note on the Efficiency of Seemingly Unrelated Regression,” The American Statistician, 42: 137-139. Boot, J. and G. de Witt (1960), “Investment Demand: An Empirical Contribution to the Aggregation Problem,” International Economic Review, 1: 3-30.


250

Chapter 10: Seemingly Unrelated Regressions

Breusch, T.S. and A.R. Pagan (1980), “The Lagrange Multiplier Test and Its Applications to Model Specification in Econometrics,” Review of Economic Studies, 47: 239-253. Conniffe, D. (1982), “A Note on Seemingly Unrelated Regressions,” Econometrica, 50: 229-233. Dwivedi, T.D. and V.K. Srivastava (1978), “Optimality of Least Squares in the Seemingly Unrelated Regression Equations Model,” Journal of Econometrics, 7: 391-395. Fiebig, D.G. (2001), “Seemingly Unrelated Regression,” Chapter 5 in Baltagi, B.H. (ed.), A Companion to Theoretical Econometrics (Blackwell: Massachusetts). Grunfeld, Y. (1958), “The Determinants of Corporate Investment,” unpublished Ph.D. dissertation (University of Chicago: Chicago, IL). Hocking, R.R. and W.B. Smith (1968), “Estimation of Parameters in the Multivariate Normal Distribution with Missing Observations,” Journal of the American Statistical Association, 63: 154-173. Hwang, H.S. (1990), “Estimation of Linear SUR Model With Unequal Numbers of Observations,” Review of Economics and Statistics, 72: 510-515. Im, Eric Iksoon (1994), “Trace Minimization of Singular Systems With Cross-Equation Restrictions,” Econometric Theory, Solution 93.2.4, 10: 450. Kmenta, J. and R. Gilbert (1968), “Small Sample Properties of Alternative Estimators of Seemingly Unrelated Regressions,” Journal of the American Statistical Association, 63: 1180-1200. Milliken, G.A. and M. Albohali (1984), “On Necessary and Sufficient Conditions for Ordinary Least Squares Estimators to be Best Linear Unbiased Estimators,” The American Statistician, 38: 298299. Oberhofer, W. and J. Kmenta (1974), “A General Procedure for Obtaining Maximum Likelihood Estimates in Generalized Regression Models,” Econometrica, 42: 579-590. Qian, H. (1998), “Equivalence of LR Test and Hausman Test,” Econometric Theory, Problem 98.1.3, 14: 151. Revankar, N.S. (1974), “Some Finite Sample Results in the Context of Two Seemingly Unrelated Regression Equations,” Journal of the American Statistical Association, 71: 183-188. Rossi, P.E. (1989), “The ET Interview: Professor Arnold Zellner,” Econometric Theory, 5: 287-317. Schmidt, P. (1977), “Estimation of Seemingly Unrelated Regressions With Unequal Numbers of Observations,” Journal of Econometrics, 5: 365-377. Schmidt, P. (1978), “A Note on the Estimation of Seemingly Unrelated Regression Systems,” Journal of Econometrics, 7: 259-261. Sentana, E. (1997), “Estimation of a Triangular, Seemingly Unrelated, Regression System by OLS,” Econometric Theory, Problem 97.2.2, 13: 463. Srivastava, V.K. and T.D. Dwivedi (1979), “Estimation of Seemingly Unrelated Regression Equations: A Brief Survey,” Journal of Econometrics, 10: 15-32. Srivastava, V.K. and D.E.A. Giles (1987), Seemingly Unrelated Regression Equations Models: Estimation and Inference (Marcel Dekker: New York). Srivastava, J.N. and M.K. Zaatar (1973), “Monte Carlo Comparison of Four Estimators of Dispersion Matrix of a Bivariate Normal Population, Using Incomplete Data,” Journal of the American Statistical Association, 68: 180-183.


References

251

Telser, L. (1964), “Iterative Estimation of a Set of Linear Regression Equations,” Journal of the American Statistical Association, 59: 845-862. Wilks, S.S. (1932), “Moments and Distributions of Estimates of Population Parameters From Fragmentary Samples,” Annals of Mathematical Statistics, 3: 167-195. Zellner, A. (1962), “An Efficient Method of Estimating Seemingly Unrelated Regressions and Tests for Aggregation Bias,” Journal of the American Statistical Association, 57: 348-368.


CHAPTER 11

Simultaneous Equations Model 11.1

Introduction

Economists formulate models for consumption, production, investment, money demand and money supply, labor demand and labor supply to attempt to explain the workings of the economy. These behavioral equations are estimated equation by equation or jointly as a system of equations. These are known as simultaneous equations models. Much of today’s econometrics have been influenced and shaped by a group of economists and econometricians known as the Cowles Commission who worked together at the University of Chicago in the late 1940’s, see Chapter 1. Simultaneous equations models had their genesis in economics during that period. Haavelmo’s (1944) work emphasized the use of the probability approach to formulating econometric models. Koopmans and Marschak (1950) and Koopmans and Hood (1953) in two influential Cowles Commission monographs provided the appropriate statistical procedures for handling simultaneous equations models. In this chapter, we first give simple examples of simultaneous equations models and show why the least squares estimator is no longer appropriate. Next, we discuss the important problem of identification and give a simple necessary but not sufficient condition that helps check whether a specific equation is identified. Sections 11.2 and 11.3 give the estimation of a single and a system of equations using instrumental variable procedures. Section 11.4 gives a test of over-identification restrictions whereas, section 11.5 gives a Hausman specification test. Section 11.6 concludes with an empirical example. The Appendix revisits the identification problem and gives a necessary and sufficient condition for identification. 11.1.1

Simultaneous Bias

Example 1: Consider a simple Keynesian model with no government Ct = α + βYt + ut

t = 1, 2, . . . , T

Y t = C t + It

(11.1) (11.2)

where Ct denotes consumption, Yt denotes disposable income, and It denotes autonomous investment. This is a system of two simultaneous equations, also known as structural equations with the second equation being an identity. The first equation can be estimated by OLS giving T T 2  β OLS = t=1 yt ct / t=1 yt

 ¯ and α  OLS = C¯ − β OLS Y

(11.3)

 with yt and ct denoting Yt and Ct in deviation form, i.e., yt = Yt − Y¯ , and Y¯ = Tt=1 Yt /T . Since It is autonomous, it is an exogenous variable determined outside the system, whereas Ct and Yt are endogenous variables determined by the system. Let us solve for Yt and Ct in terms of the constant and It . The resulting two equations are known as the reduced form equations Ct = α/(1 − β) + βIt (1 − β) + ut /(1 − β) Yt = α/(1 − β) + It /(1 − β) + ut /(1 − β)

(11.4) (11.5)


254

Chapter 11: Simultaneous Equations Model

These equations express each endogenous variable in terms of exogenous variables and the error terms. Note that both Yt and Ct are a function of ut , and hence both are correlated with ut . In fact, Yt − E(Yt ) = ut /(1 − β), and cov(Yt , ut ) = E[(Yt − E(Yt ))ut ] = σ 2u /(1 − β) ≥ 0 if

0≤β≤1

(11.6)

This holds because ut ∼ (0, σ 2u ) and It is exogenous and independent of the error term. Equation (11.6) shows that the right hand side regressor in (11.1) is correlated with the error term. This causes the OLS estimates to be biased and inconsistent. In fact, from (11.1), ¯) ct = Ct − C¯ = βyt + (ut − u and substituting this expression in (11.3), we get T T 2  β OLS = β + t=1 yt ut / t=1 yt

(11.7)

 OLS ) = β, since the expected value of the second term is not From (11.7), it is clear that E(β necessarily zero. Also, using (11.5) one gets yt = Yt − Y¯ = [it + (ut − u ¯)]/(1 − β)   where it = It − I¯ and I¯ = Tt=1 It /T . Defining myy = Tt=1 yt2 /T , we get myy = (mii + 2miu + muu )/(1 − β)2

where mii =

T

2 t=1 it /T ,

miu =

T

t=1 it (ut

−u ¯)/T and muu =

myu = (miu + muu )/(1 − β)

T

t=1 (ut

(11.8)

−u ¯)2 /T . Also, (11.9)

Using the fact that plim miu = 0 and plim muu = σ 2u , we get 2 2  plim β OLS = β + plim (myu /myy ) = β + [σ u (1 − β)/(plim mii + σ u )]

 which shows that β OLS overstates β if 0 ≤ β ≤ 1.

Example 2: Consider a simple demand and supply model Qdt = α + βPt + u1t

(11.10)

Qst Qdt

(11.11)

= γ + δPt + u2t =

Qst

= Qt

t = 1, 2, . . . , T

(11.12)

Substituting the equilibrium condition (11.12) in (11.10) and (11.11), we get Qt = α + βPt + u1t Qt = γ + δPt + u2t

(11.13) t = 1, 2, . . . , T

(11.14)

For the demand equation (11.13), the sign of β is expected to be negative, while for the supply equation (11.14), the sign of δ is expected to be positive. However, we only observe one equilibrium pair (Qt , Pt ) and these are not labeled demand or supply quantities and prices. When we run the OLS regression of Qt on Pt we do not know what we are estimating, demand or supply? In fact, any linear combination of (11.13) and (11.14) looks exactly like (11.13) or (11.14). It


11.1

Introduction

255

will have a constant, Price, and a disturbance term in it. Since demand or supply cannot be distinguished from this ‘mongrel’ we have what is known as an identification problem. If the demand equation (or the supply equation) looked different from this mongrel, then this particular equation would be identified. More on this later. For now let us examine the properties of the OLS estimates of the demand equation. It is well known that  OLS = T qt pt / T p2 = β + T pt (u1t − u ¯1 )/ Tt=1 p2t (11.15) β t=1 t=1 t t=1

¯ This estimator is where qt and pt denote Qt and Pt in deviation form, i.e., qt = Qt − Q. unbiased depending on whether the last term in (11.15) has zero expectations. In order to find this expectation we solve the structural equations in (11.13) and (11.14) for Qt and Pt Qt = (αδ − γβ)/(δ − β) + (δu1t − βu2t )/(δ − β) Pt = (α − γ)/(δ − β) + (u1t − u2t )/(δ − β)

(11.16) (11.17)

(11.16) and (11.17) are known as the reduced form equations. Note that both Qt and Pt are functions of both errors u1 and u2 . Hence, Pt is correlated with u1t . In fact, pt = (u1t − u ¯1 )/(δ − β) − (u2t − u ¯2 )/(δ − β)

(11.18)

and plim plim

T

t=1 pt (u1t

T

2 t=1 pt /T

−u ¯1 )/T = (σ 11 − σ 12 )/(δ − β)

(11.19)

= (σ 11 + σ 22 − 2σ 12 )/(δ − β)2

(11.20)

where σ ij = cov(uit , ujt ) for i, j = 1, 2; and t = 1, . . . , T . Hence, from (11.15)  plim β OLS = β + (σ 11 − σ 12 )(δ − β)/(σ 11 + σ 22 − 2σ 12 )

(11.21)

 and the last term is not necessarily zero, implying that β OLS is not consistent for β. Similarly, one can show that the OLS estimator for δ is not consistent, see problem 1. This simultaneous bias is once again due to the correlation of the right hand side variable (price) with the error term u1 . This correlation could be due to the fact that Pt is a function of u2t , from (11.17), and u2t and u1t are correlated, making Pt correlated with u1t . Alternatively, Pt is a function of Qt , from (11.13) or (11.14), and Qt is a function of u1t , from (11.13), making Pt a function of u1t . Intuitively, if a shock in demand (i.e., a change in u1t ) shifts the demand curve, the new intersection of demand and supply determines a new equilibrium price and quantity. This new price is therefore, affected by the change in u1t , and is correlated with it. In general, whenever a right hand side variable is correlated with the error term, the OLS estimates are biased and inconsistent. We refer to this as an endogeneity problem. Recall, Figure 3 of Chapter 3 with cov(Pt , u1t ) > 0. This shows that Pt ’s above their mean are on the average associated with u1t ’s above their mean, (i.e., u1t > 0). This implies that the quantity Qt associated with this particular Pt is on the average above the true line (α + βPt ). This is true for all observations to the right of E(Pt ). Similarly, any Pt to the left of E(Pt ) is on the average associated with a u1t below its mean, (i.e., u1t < 0). This implies that quantities associated with prices below their mean E(Pt ) are on the average data points that lie below the true line. With this observed data, the estimated line using OLS will always be biased. In this case, the intercept estimate is biased downwards, whereas the slope estimate is biased upwards. This bias does not


256

Chapter 11: Simultaneous Equations Model

disappear with more data, as any new observation will on the average be either above the true line if Pt > E(Pt ) or below the line if Pt < E(Pt ). Hence, these OLS estimates are inconsistent. Deaton (1997, p. 95) has a nice discussion of endogeneity problems in development economics. One important example pertains to farm size and farm productivity. Empirical studies using OLS have found an inverse relationship between productivity as measured by log(Output/Acre) and farm size as measured by (Acreage). This seems counter-intuitive as it suggests that smaller farms are more productive than larger farms. Economic explanations of this phenomenon include the observation that hired labor (which is typically used on large farms) is of lower quality than family labor (which is typically used on small farms). The latter needs less monitoring and can be entrusted with valuable animals and machinery. Another explanation is that this phenomenon is an optimal response by small farmers to uncertainty. It could also be a sign of inefficiency as farmers work too much on their own farms pushing their marginal productivity below market wage. How could this be an endogeneity problem? After all, the amount of land is outside the control of the farmer. This is true, but that does not mean that acreage is uncorrelated with the disturbance term. After all, size is unlikely to be independent of the quality of land. “Desert farms that are used for low-intensity animal grazing are typically larger than garden farms, where the land is rich and output/acre is high.” In this case, land quality is negatively correlated with land size. It takes more acres to sustain a cow in West Texas than in less arid areas. This negative correlation between acres, the explanatory variable and quality of land which is an omitted variable included in the error term introduces endogeneity. This in turn results in downward bias of the OLS estimate of acreage on productivity. Endogeneity can also be caused by sample selection. Gronau (1973) observed that women with small children had higher wages than women with no children. An economic explanation is that women with children have higher reservation wages and as a result fewer of them work. Of those that work, their observed wages are higher than those without children. The endogeneity works through the unobserved component in the working women’s wage that induces her to work. This is positively correlated with the number of children she has and therefore introduces upward biases in the OLS estimate of the effect of the number of children on wages. 11.1.2

The Identification Problem

In general, we can think of any structural equation, say the first, as having one left hand side endogenous variable y1 , g1 right hand side endogenous variables, and k1 right hand side exogenous variables. The right hand side endogenous variables are correlated with the error term rendering OLS on this equation biased and inconsistent. Normally, for each endogenous variable, there exists a corresponding structural equation explaining its behavior in the model. We say that a system of simultaneous equations is complete if there are as many endogenous variables as there are equations. To correct for the simultaneous bias we need to replace the right hand side endogenous variables in this equation by variables which are highly correlated with the ones they are replacing but not correlated with the error term. Using the method of instrumental variable estimation, discussed below, we will see that these variables turn out to be the predictors obtained by regressing each right hand side endogenous variable on a subset of all the exogenous variables in the system. Let us assume that there are K exogenous variables in the simultaneous system. What set of exogenous variables should we use that would lead to consistent estimates of this structural equation? A search for the minimum set needed for consistency leads us to the order condition for identification.


11.1

Introduction

257

The Order Condition for Identification: A necessary condition for identification of any structural equation is that the number of excluded exogenous variables from this equation are greater than or equal to the number of right hand side included endogenous variables. Let K be the number of exogenous variables in the system, then this condition requires k2 ≥ g1 , where k2 = K − k1 . Let us consider the demand and supply equations given in (11.13) and (11.14) but assume that the supply equation has in it an extra variable Wt denoting weather conditions. In this case the demand equation has one right hand side endogenous variable Pt , i.e., g1 = 1 and one excluded exogenous variable Wt , making k2 = 1. Since k2 ≥ g1 , this order condition is satisfied, in other words, based on the order condition alone we cannot conclude that the demand equation is unidentified. The supply equation, however, has g1 = 1 and k2 = 0, making this equation unidentified, since it does not satisfy the order condition for identification. Note that this condition is only necessary but not sufficient for identification. In other words, it is useful only if it is not satisfied, in which case the equation in question is not identified. Note that any linear combination of the new supply and demand equations would have a constant, price and weather. This looks like the supply equation but not like demand. This is why the supply equation is not identified. In order to prove once and for all whether the demand equation is identified, we need the rank condition for identification and this will be discussed in details in the Appendix to this chapter. Adding a third variable to the supply equation like the amount of fertilizer used Ft will not help the supply equation any, since a linear combination of supply and demand will still look like supply. However, it does help the identification of the demand equation. Denote by ℓ = k2 − g1 , the degree of over-identification. In (11.13) and (11.14) both equations are unidentified (or under-identified) with ℓ = −1. When Wt is added to the supply equation, ℓ = 0 for the demand equation, and it is just-identified. When both Wt and Ft are included in the supply equation, ℓ = 1 and the demand equation is over-identified. Without the use of matrices, we can describe a two-stage least squares method that will estimate the demand equation consistently. First, we run the right hand side endogenous variable Pt on a constant and Wt and get Pt , then replace Pt in the demand equation with Pt and perform this second stage regression. In other words, the first step regression is Pt = π 11 + π 12 Wt + vt with vt = Pt − Pt satisfying the OLS normal equations stage regression is with

Qt = α + β Pt + ǫt

T

ǫt t=1 

=

T

(11.22) T

t t=1 v

=

T

t Wt t=1 v

 = 0. Using (11.13) and (11.23), we can write

ǫt Pt t=1 

ǫt = β(Pt − Pt ) + u1t = β vt + u1t

= 0. The second

(11.23)

(11.24)

T T T T T   t = so that t=1 v t=1 u1t Pt using the fact that t=1 ǫt = t=1 u1t and t=1 ǫt Pt = T t = 0. So the new error ǫt behaves as the original disturbance u1t . However, our P v  t t=1 right hand side variable is now Pt which is independent of u1t since it is a linear combination of exogenous variables only. We essentially decomposed Pt into two parts, the first part Pt is a linear combination of exogenous variables and therefore, independent of the u1t ’s. The second


258

Chapter 11: Simultaneous Equations Model

part is  vt which is correlated with u1t . In fact, this is the source of simultaneous bias. The two parts Pt and vt are orthogonal to each other by construction. Hence when the vt ’s become part of the new error ǫt , they are orthogonal to the new regressor Pt . Furthermore, Pt is also independent of u1t . Why would this procedure not work on the estimation of (11.13) if the model is given by equations (11.13) and (11.14). The answer is that in (11.22) we will only have a constant, and no Wt . When we try to run the second-stage regression in (11.23) the regression will fail because of perfect multicollinearity between the constant and Pt . This will happen whenever the order condition is not satisfied and the equation is not identified, see Kelejian and Oates (1989). Hence, in order for it to succeed in the second stage we need at least one excluded exogenous variable from the demand equation that is in the supply equation, i.e., variables like Wt or Ft . Therefore, whenever the second-stage regression fails because of perfect multicollinearity between the right hand side regressors, this implies that the order condition of identification is not satisfied. In general, if we are given an equation like y1 = α12 y2 + β 11 X1 + β 12 X2 + u1

(11.25)

the order condition requires the existence of at least one exogenous variable excluded from (11.25), say X3 . These extra exogenous variables like X3 usually appear in other equations of our simultaneous equation model. In the first step regression we run y2 = π 21 X1 + π 22 X2 + π 23 X3 + v2 with the OLS residuals v2 satisfying T T 2t X2t = 0; 2t X1t = 0; t=1 v t=1 v

(11.26) T

and in the second step, we run the regression

2t X3t t=1 v

y1 = α12 y2 + β 11 X1 + β 12 X2 + ǫ1

=0

(11.27)

(11.28)

where ǫ1 = α12 (y2 − y2 ) + u1 = α12 v2 + u1 . This regression will lead to consistent estimates, because  T T T y2t ǫ1t = Tt=1 y2t u1t ; t=1 X1t ǫ1t = t=1 X1t u1t ; t=1  (11.29) T T X ǫ = X u t=1 2t 1t t=1 2t 1t

and u1t is independent of the exogenous variables. In order to solve for 3 structural  parameters α12 , β 11 and β 12 one needs three linearly independent OLS normal equations. Tt=1 y2tǫ1t = 0 is a new piece of information provided y2 is regressed on at least one extra variable besides  X1 and X2 . Otherwise, Tt=1 X1tǫ1t = Tt=1 X2tǫ1t = 0 are the only two linearly independent normal equations in three structural parameters. What happens if there is another right hand side endogenous variable, say y3 ? In that case (11.25) becomes y1 = α12 y2 + α13 y3 + β 11 X1 + β 12 X2 + u1

(11.30)

Now we need at least two exogenous variables that are excluded from (11.30) for the order condition to be satisfied, and the second stage regression to run. Otherwise, we will have less


11.2

Single Equation Estimation: Two-Stage Least Squares

259

linearly independent equations than there are structural parameters to estimate, and the second stage regression will fail. Also, y2 and y3 should be regressed on the same set of exogenous variables. Furthermore, this set of second-stage regressors should always include the right hand side exogenous variables of (11.30). These two conditions will ensure consistency of the estimates. Let X3 and X4 be the excluded exogenous variables from (11.30). Our first step regression would regress y2 and y3 on X1 , X2 , X3 and X4 to get y2 and y3 , respectively. The second stage regression would regress y1 on y2 , y3 , X1 and X2 . From the first step regressions we have y2 = y2 + v2

and y3 = y3 + v3

(11.31)

where y2 and y3 are linear combinations of the X’s, and v2 and v3 are the residuals. The second stage regression has the following normal equations T    2tǫ1t = Tt=1 y3tǫ1t = Tt=1 X1tǫ1t = Tt=1 X2tǫ1t = 0 (11.32) t=1 y

where ǫ1 denotes the residuals from the second stage regression. In fact

ǫ1 = α12 v2 + α13 v3 + u1 (11.33)     Now Tt=1 ǫ1t y2t = Tt=1 u1t y2t because Tt=1 v2t y2t = Tt=1 v3t y2t = 0. The latter holds because y2 , the predictor, is orthogonal to  v2 , the residual. Also, y2 is orthogonal to v3 if y2 is regressed on aset of X’s that are a subset of the regressors included in the first step regression of y3 . Similarly, T T ǫ y  = u y  if y is regressed on a set of exogenous variables that are a subset of 1t 3t 1t 3t 3 t=1 t=1 the X’s included in the first step regression of y2 . Combining these two conditions leads to the following fact: y2 and y3 have to be regressed on the same set of exogenous variables for the composite error term to behave like the original error. Furthermore these exogenous variables should include the included X’s on the right hand side of theequation to be estimated,  i.e., X1  and X2 , otherwise, Tt=1 ǫ1t X1t is not necessarily equal to Tt=1 u1t X1t , because Tt=1 v2t X1t  or Tt=1 v3t X1t are not necessarily zero. For further analysis along these lines, see problem 2.

11.2

Single Equation Estimation: Two-Stage Least Squares

In matrix form, we can write the first structural equation as y1 = Y1 α1 + X1 β 1 + u1 = Z1 δ 1 + u1

(11.34)

where y1 and u1 are (T × 1), Y1 denotes the right hand side endogenous variables which is (T × g1 ) and X1 is the set of right hand side included exogenous variables which is (T × k1 ), α1 is of dimension g1 and β 1 is of dimension k1 . Z1 = [Y1 , X1 ] and δ ′1 = (α′1 , β ′1 ). We require the existence of excluded exogenous variables, from (11.34), call them X2 , enough to identify this equation. These excluded exogenous variables appear in the other equations in the simultaneous model. Let the set of all exogenous variables be X = [X1 , X2 ] where X is of dimension (T × k). For the order condition to be satisfied for equation (11.34) we must have (k − k1 ) ≥ g1 . If all the exogenous variables in the system are included in the first step regression, i.e., Y1 is regressed on X to get Y1 , the resulting second stage least squares estimator obtained from regressing y1 on Y1 and X1 is called two-stage least squares (2SLS). This method was proposed independently by Basmann (1957) and Theil (1953). In matrix form Y1 = PX Y1 is the predictor of the right


260

Chapter 11: Simultaneous Equations Model

hand side endogenous variables, where PX is the projection matrix X(X ′ X)−1 X ′ . Replacing Y1 by Y1 in (11.34), we get y1 = Y1 α1 + X1 β 1 + w1 = Z1 δ 1 + w1

(11.35)

 1 )−1 Z′ y1 = (Z ′ PX Z1 )−1 Z ′ PX y1 δ 1,2SLS = (Z1′ Z 1 1 1

(11.36)

W ′ y1 = W ′ Z1 δ1 + W ′ u1

(11.37)

1 = [Y1 , X1 ] and w1 = u1 + (Y1 − Y1 )α1 . Running OLS on (11.35) one gets where Z

1 = PX Z1 and the fact that PX is where the second equality follows from the fact that Z idempotent. The former equality holds because PX X = X, hence PX X1 = X1 , and PX Y1 = Y1 . If there is only one right hand side endogenous variable, running the first-stage regression y2 on X1 and X2 and testing that the coefficients of X2 are all zero against the hypothesis that at least one of these coefficients is different from zero is a test for rank identification. In case of several right hand side endogenous variables, things get complicated, see Cragg and Donald (1996), but one can still run the first-stage regressions for each right hand side endogenous variable to make sure that at least one element of X2 is significantly different from zero.1 This is not sufficient for the rank condition but it is a good diagnostic for whether the rank condition fails. If we fail to meet this requirement we should question our 2SLS estimator. Two-stage least squares can also be thought of as a simple instrumental variables estimator 1 = [Y1 , X1 ]. Recall that Y1 is correlated with u1 , rendering with the set of instruments W = Z OLS inconsistent. The idea of simple instrumental variables is to find a set of instruments, say W for Z1 with the following properties: (1) plim W ′ u1 /T = 0, the instruments have to be exogenous, i.e., uncorrelated with the error term, otherwise this defeats the purpose of the instruments and result in inconsistent estimates. (2) plim W ′ W/T = Qw = 0, where Qw is finite and positive definite, the W ′ s should not be perfectly multicollinear. (3) W should be highly correlated with Z1 , i.e., the instruments should be highly relevant, not weak instruments as we will explain shortly. In fact, plim W ′ Z1 /T should be finite and of full rank (k1 + g1 ). Premultiplying (11.34) by W ′ , we get

In this case, W = Z1 is of the same dimension as Z1 , and since plim W ′ Z1 /T is square and of full rank (k1 + g1 ), the simple instrumental variable (IV) estimator of δ 1 becomes  δ 1,IV = (W ′ Z1 )−1 W ′ y1 = δ 1 + (W ′ Z1 )−1 W ′ u1

(11.38)

with plim  δ 1,IV = δ 1 which follows from (11.37) and the fact that plim W ′ u1 /T = 0.

Digression: In the general linear model, y = Xβ + u, X is the set of instruments for X. Premultiplying by X ′ we get X ′ y = X ′ Xβ + X ′ u and using the fact that plim X ′ u/T = 0, one gets  OLS . IV = (X ′ X)−1 X ′ y = β β

This estimator is consistent as long as X and u are uncorrelated. In the simultaneous equation model for the first structural equation given in (11.34), the right hand side regressors Z1 include endogenous variables Y1 that are correlated with u1 . Therefore OLS on (11.34) will lead to


11.2

Single Equation Estimation: Two-Stage Least Squares

261

inconsistent estimates, since the matrix of instruments W = Z1 , and Z1 is correlated with u1 . In fact,  δ 1,OLS = (Z1′ Z1 )−1 Z1′ y1 = δ 1 + (Z1′ Z1 )−1 Z1′ u1

with plim  δ 1,OLS = δ 1 since plim Z1′ u1 /T = 0. Denote by e1,OLS = y1 − Z1 δ 1,OLS as the OLS residuals on the first structural equation, then plim s21 =

e′1,OLS e1,OLS

T − (g1 + k1 )

= plim

u′ Z1 (Z1′ Z1 )−1 Z1′ u1 u′1 P¯Z1 u1 = σ 11 − plim 1 ≤ σ 11 , T − (g1 + k1 ) T − (g1 + k1 )

since the last term is positive. Only if plim Z1′ u1 /T is zero will plim s21 = σ 11 , otherwise it is smaller. OLS fits very well, it minimizes (y1 − Z1 δ1 )′ (y1 − Z1 δ 1 ). Since Z1 and u1 are correlated, OLS attributes part of the variation in y1 that is due to u1 incorrectly to the regressor Z1 . Both the simple IV and OLS estimators can be interpreted as method of moments estimators. These were discussed in Chapter 2. For OLS, the population moment conditions are given  = 0. by E(X ′ u) = 0 and the corresponding sample moment conditions yield X ′ (y − X β)/T  results in β  Solving for β . Similarly, the population moment conditions for the simple IV OLS estimator in (11.37) are E(W ′ u1 ) = 0 and the corresponding sample moment conditions yield W ′ (y1 − Z1 δ 1 )/T = 0. Solving for  δ1 results in  δ 1,IV given in (11.38).  If W = [Y1 , X1 ], then (11.38) results in  δ 1,IV =

Y1′ Y1 Y1′ X1 X1′ Y1 X1′ X1

−1

Y1′ y1 X1′ y1

which is the same as (11.36) −1

 Y1′ Y1 Y1′ X1 Y1′ y1  δ 1,2SLS = X1′ y1 X1′ Y1 X1′ X1

(11.39)

(11.40)

provided Y1′ Y1 = Y1′ Y1 , and X1′ Y1 = X1′ Y1 . The latter conditions hold because Y1 = PX Y1 , and PX X1 = X1 . In general, let X ∗ be our set of first stage regressors. An IV estimator with Y1∗ = PX ∗ Y1 , i.e., with every right hand side y regressed on the same set of regressors X ∗ , will satisfy Y1∗′ Y1∗ = Y1∗′ PX ∗ Y1 = Y1∗′ Y1

In addition, for X1′ Y1∗ to equal X1′ Y1 , X1 has to be a subset of the regressors in X ∗ . Therefore X ∗ should include X1 and at least as many X’s from X2 as is required for identification, i.e., (at least g1 of the X’s from X2 ). In this case, the IV estimator using W ∗ = [Y1∗ , X1 ] will result in the same estimator as that obtained by a two stage regression where in the first step Y1∗ is obtained by regressing Y1 on X ∗ , and in the second step y1 is regressed on W ∗ . Note that these are the same conditions required for consistency of an IV estimator. Note also, that if this equation is just-identified, then there is exactly g1 of the X’s excluded from that equation. In other words, X2 is of dimension (T × g1 ), and X ∗ = X is of dimension T × (g1 + k1 ). Problem 3 shows that 2SLS in this case reduces to an IV estimator with W = X, i.e.  δ 1,2SLS =  δ 1,IV = (X ′ Z1 )−1 X ′ y1

(11.41)


262

Chapter 11: Simultaneous Equations Model

Note that if the first equation is over-identified, then X ′ Z1 is not square and (11.41) cannot be computed. Rather than having W , the matrix of instruments, be of exactly the same dimension as Z1 which is required for the expression in (11.38), one can define a generalized instrumental variable in terms of a general matrix W of dimension T × ℓ where ℓ ≥ g1 + k1 . The latter condition is the order condition for identification. In this case,  δ1,IV is obtained as GLS on (11.37). Using the fact that plim W ′ u1 u′1 W/T = σ 11 plim W ′ W/T,

one gets  δ 1,IV = (Z1′ PW Z1 )−1 Z1′ PW y1 = δ 1 + (Z1′ PW Z1 )−1 Z1′ PW u1

with plim  δ 1,IV = δ 1 and limiting covariance matrix σ 11 plim (Z1′ PW Z1 /T )−1 . Therefore, 2SLS can be obtained as a generalized instrumental variable estimator with W = X. This also means that 2SLS of δ 1 can be obtained as GLS on (11.34) after premultiplication by X ′ , see problem 4. Note that GLS on (11.37) minimizes (y1 − Z1 δ 1 )′ PW (y1 − Z1 δ 1 ) which yields the first-order conditions Z1′ PW (y1 − Z1 δ1,IV ) = 0

the solution of which is  δ 1,IV = (Z1′ PW Z1 )−1 Z1′ PW y1 . It can also be shown that 2SLS and the generalized instrumental variables estimators are special cases of a Generalized Method of Moments (GMM) estimator considered by Hansen (1982). See Davidson and MacKinnon (1993) and Hall (1993) for an introduction to GMM. For the matrix Z1′ PW Z1 to be of full rank and invertible, a necessary condition is that W must be of full rank ℓ ≥ (g1 + k1 ). This is in fact, the order condition of identification. If ℓ = g1 + k1 , then this equation is just-identified. Also, W ′ Z1 is square and nonsingular. Problem 10 asks the reader to verify that the generalized instrumental variable estimator reduces to the simple instrumental variable estimator given in (11.38). Also, under just-identification the minimized value of the criterion function is zero. One of the biggest problems with IV estimation is the choice of the instrumental variables W . We have listed some necessary conditions for this set of instruments to yield consistent estimators of the structural coefficients. However, different choices by different researchers may yield different estimates in finite samples. Using more instruments will yield more efficient IV estimation. Let W1 and W2 be two sets of IV’s with W1 being spanned by the space of W2 . In this case, PW2 W1 = W1 and therefore, PW2 PW1 = PW1 . For the corresponding IV estimators  δ 1,Wi = (Z1′ PWi Z1 )−1 Z1′ PWi y1

for i = 1, 2

are both consistent for δ 1 as long as plim Wi′ u1 /T = 0 and have asymptotic covariance matrices σ 11 plim (Z1′ PWi Z1 /T )−1 Note that  δ1,W2 is at least as efficient as  δ 1,W1 , if the difference in their asymptotic covariance matrices is positive semi-definite, i.e., if

Z1′ PW2 Z1 −1 Z1′ PW1 Z1 −1 − σ 11 plim σ 11 plim T T


11.2

Single Equation Estimation: Two-Stage Least Squares

263

is p.s.d. This holds, if Z1′ PW2 Z1 − Z1′ PW1 Z1 is p.s.d. This last condition holds since PW2 − PW1 is idempotent. Problem 11 asks the reader to verify this result.  δ1,W2 is more efficient than  δ 1,W1 since W2 explains Z1 at least as well as W1 . This seems to suggest that one should use as many instruments as possible. If T is large this is a good strategy. But, if T is finite, there will be a trade-off between this gain in asymptotic efficiency and the introduction of more finite sample bias in our IV estimator. In fact, the more instruments we use, the more will Y1 resemble Y1 and the more bias is introduced in this second stage regression. The extreme case where Y1 is perfectly predicted by Y1 returns us to OLS which we know is biased. On the other hand, if our set of instruments have little ability in predicting Y1 , then the resulting instrumental variable estimator will be inefficient and its asymptotic distribution will not resemble its finite sample distribution, see Nelson and Startz (1990). If the number of instruments is fixed √ and the coefficients of the instruments in the first stage regression go to zero at the rate 1 T , indicating weak correlation, Staiger and Stock (1997) find that even as T increases, IV estimation is not consistent and has a nonstandard asymptotic distribution. Bound et al. (1995) recommend reporting the R2 or the F -statistic of the first stage regression as a useful indicator of the quality of IV estimates. Instrumental variables are important for obtaining consistent estimates when endogeneity is suspected. However, invalid instruments can produce meaningless results. How do we know whether our instruments are valid? Stock and Watson (2003) draw an analogy between a relevant instrument and a large sample. The more relevant the instrument, i.e., the more the variation in the right hand side endogenous variable that is explained by this instrument, the more accurate the resulting estimator. This is similar to the observation that the larger the sample size, the more accurate the estimator. They argue that the instruments should not just be relevant, but highly relevant if the normal distribution is to provide a good approximation to the sampling distribution of 2SLS. Weak instruments explain little of the variation in the right hand side endogenous variable they are instrumenting. This renders the normal distribution as a poor approximation to the sampling distribution of 2SLS, even if the sample size is large. Stock and Watson (2003) suggest a simple rule of thumb to check for weak instruments. If there is one right hand side endogenous variable, the first-stage regression can test for the significance of the excluded exogenous variables (or instruments) using an F-statistic. This first-stage F -statistic should be larger than 10.2 Stock and Watson (2003) suggest that a first-stage F -statistic less than 10 indicates weak instruments which casts doubt on the validity of 2SLS, since with weak instruments, 2SLS will be biased even in large samples and the corresponding t-statistics and confidence intervals will be unreliable. Finding weak instruments, one can search for additional stronger instruments, or use alternative estimators than 2SLS which are less sensitive to weak instruments like LIML. Deaton (1997, p. 112) argues that it is difficult to find instruments that are exogenous while at the same time highly correlated with the endogenous variables they are instrumenting. He argues that it is easy to generate 2SLS estimates that are different from OLS but much harder to make the case that these 2SLS estimates are necessarily better than OLS. “Credible identification and estimation of structural equations almost always requires real creativity, and creativity cannot be reduced to a formula.” Stock and Watson (2003, p. 371) show that for the case of a single right hand side endogenous variable with no included exogenous variables and one weak instrument, the distribution of the 2SLS estimators is nonnormal even for large samples, with the mean of the sampling distribution of the 2SLS estimator approximately equal to the true coefficient plus the asymptotic bias of the OLS estimator divided


264

Chapter 11: Simultaneous Equations Model

by (E(F ) − 1) where F is the first-stage F -statistic. If E(F ) = 10, then the large sample bias of 2SLS is (1/9) that of the large sample bias of OLS. They argue that this rule of thumb is an acceptable cutoff for most empirical applications. 2SLS is a single equation estimator. The focus is on a particular equation. [y1 , Y1 , X1 ] is specified and therefore all that is needed to perform 2SLS is the matrix X of all exogenous variables in the system. If a researcher is interested in a particular behavioral economic relationship which may be a part of a big model consisting of several equations, one need not specify the whole model to perform 2SLS on that equation, all that is needed is the matrix of all exogenous variables in that system. Empirical studies involving one structural equation, specify which right hand side variables are endogenous and proceed by estimating this equation via an IV procedure that usually includes all the feasible exogenous variables available to the researcher. If this set of exogenous variables does not include all the X’s in the system, this estimation method is not 2SLS. However, it is a consistent IV method which we will call feasible 2SLS. Substituting (11.34) in (11.36), we get  δ 1,2SLS = δ 1 + (Z1′ PX Z1 )−1 Z1′ PX u1

(11.42)

with plim  δ 1,2SLS = δ 1 and an asymptotic variance covariance matrix given by σ 11 plim (Z1′ PX Z1 /T )−1 . σ 11 is estimated from the 2SLS residuals u 1 = y1 − Z1 δ 1,2SLS , by comput′ ing s11 = u 1 u 1 /(T − g1 − k1 ). It is important to emphasize that s11 is obtained from the 2SLS residuals of the original equation (11.34), not (11.35). In other words, s11 is not the mean squared error (i.e., s2 ) of the second stage regression given in (11.35). The latter regression has Y1 in it and not Y1 . Therefore, the asymptotic variance covariance matrix of 2SLS can be esti1 )−1 . The t-statistics reported by 2SLS packages are based mated by s11 (Z1′ PX Z1 )−1 = s11 (Z1′ Z on the standard errors obtained from the square root of the diagonal elements of this matrix. These standard errors and t-statistics can be made robust for heteroskedasticity by computing 1 )−1 (Z′ diag[ 1 )−1 where u (Z1′ Z u2i ]Z1 )(Z1′ Z i denotes the i-th 2SLS residual. Wald type statistics 1 for Ho ; Rδ 1 = r based on 2SLS estimates of δ 1 can be obtained as in equation (7.41) with ′ −1     ′  −1 replacing var(β δ 1,2SLS replacing β OLS and var(δ 1,2SLS ) = s11 (Z1 Z1 ) OLS ) = s11 (X X) . This can be made robust for heteroskedasticity by using the robust variance covariance matrix of  δ 1,2SLS described above. The resulting Wald statistic is asymptotically distributed as χ2q under the null hypothesis, with q being the number of restrictions imposed by Rδ 1 = r. LM type tests for exclusion restrictions, like a subset of δ 1 set equal to zero can be performed by running the restricted 2SLS residuals on the matrix of unrestricted second stage regressors Z1 . The test statistic is given by T Ru2 where Ru2 denotes the uncentered R2 . This is asymptotically distributed as χ2q under the null hypothesis, where q is the number of coefficients in δ 1 set equal to zero. Note that it does not matter whether the exclusion restrictions are imposed on β 1 or α1 , i.e., whether the excluded variables to be tested are endogenous or exogenous. An F-test for these exclusion restrictions can be constructed based on the restricted and unrestricted residual sums of squares from the second stage regression. The denominator of this F-statistic, however, is based on the unrestricted 2SLS residual sum of squares as reported by the 2SLS package. Of course, one has to adjust the numerator and denominator by the appropriate degrees of freedom. Under the null, this is asymptotically distributed as F (q, T − (g1 + k1 )). See Wooldridge (1990) for details. Also, see the over-identification test in Section 11.5. Finite sample properties of 2SLS are model specific, see Mariano (2001) for a useful summary. One important result is that the absolute moments of positive order for 2SLS are finite up to the order of over-identification. So, for the 2SLS estimator to have a mean and variance, we


11.2

Single Equation Estimation: Two-Stage Least Squares

265

need the degree of over-identification to be at least 2. This also means that for a just-identified model, no moments for 2SLS exist. For 2SLS, the absolute bias is an increasing function of the degree of over-identification. For the case of one right hand side included endogenous regressor, like equation (11.25), the size of OLS bias relative to 2SLS gets larger, the lower the degree of over-identification, the bigger the sample size, the higher the absolute value of the correlation between the disturbances and the endogenous regressor y2 and the higher the concentration parameter μ2 . The latter is defined as μ2 = E(y2 )′ (PX − PX1 )E(y2 )/ω 2 and ω 2 = var(y2t ). In terms of MSE, larger values of μ2 and large sample size favor 2SLS over OLS. Another important single equation estimator is the Limited Information Maximum Likelihood (LIML) estimator which as the name suggests maximizes the likelihood function pertaining to the endogenous variables appearing in the estimated equation only. Excluded exogenous variables from this equation as well as the identifiability restrictions on other equations in the system are disregarded in the likelihood maximization. For details, see Anderson and Rubin (1950). LIML is invariant to the normalization choice of the dependent variable whereas 2SLS is not. This invariancy of LIML is in the spirit of a simultaneous equation model where normalization should not matter. Under just-identification 2SLS and LIML are equivalent. LIML is also known as the Least Variance Ratio (LVR) method, since the LIML estimates can be obtained by minimizing a ratio of two variances or equivalently the ratio of two residual sum of squares. Using equation (11.34), one can write y1∗ = y1 − Y1 α = X1 β 1 + u1 For a choice of α one can compute y1∗ and regress it on X1 to get the residual sum of squares RSS1 . Now regress y1∗ on X1 and X2 and compute the residual sum of squares RSS2 . Equation (11.34) states that X2 does not enter the specification of that equation. In fact, this is where our identifying restrictions come from and the excluded exogenous variables that are used as instrumental variables. If these identifying restrictions are true, adding X2 to the regression of y1∗ and X1 should lead to minimal reduction in RSS1 . Therefore, the LVR method finds the α that will minimize the ratio (RSS1 /RSS2 ). After α is estimated, β 1 is obtained from regressing y1∗ on X1 . In contrast, it can be shown that 2SLS minimizes RSS1 − RSS2 . For details, see Johnston (1984) or Mariano (2001). Estimator bias is less of a problem for LIML than 2SLS. In fact as the number of instruments increase with the sample size such that their ratio is a constant, Bekker (1994) shows that 2SLS becomes inconsistent while LIML remains consistent. Example 3: Simple Keynesian Model For the data from the Economic Report of the President, given in Table 5.1, consider the simple Keynesian model with no government Ct = α + βYt + ut

t = 1, 2, . . . , T

with Yt = Ct + It . The OLS estimates of the consumption function yield: Ct = −65.79 + 0.916 Yt (90.99) (0.009) The 2SLS estimates assuming that It is exogenous and is the only instrument available, yield Ct = 313.01 + 0.878 Yt (129.8) (0.012)


266

Chapter 11: Simultaneous Equations Model

Table 11.1 Two-Stage Least Squares Dependent Variable: Method: Sample: Included observations: Instrument list:

CONSUM Two-Stage Least Squares 1950 1993 44 I

Variable C Y

Coefficient 313.0072 0.878394

R-squared Adjusted R-squared S.E. of regression F-statistic Prob (F-statistic)

0.994620 0.994492 184.4002 4966.557 0.00000

Std. Error 129.8318 0.012464

t-Statistic 2.410867 70.47380

Mean dependent var S.D. dependent var Sum squared resid Durbin-Watson stat

Prob. 0.0204 0.0000 9250.545 2484.624 1428145 0.31203

Table 11.1 reports these 2SLS results using EViews. Note that the OLS estimate of the intercept is understated, while that of the slope estimate is overstated indicating positive correlation between Yt and the error as described in (11.6). OLS on the reduced form equations yield Ct = 2573.94 + 7.22 It (811.9) (0.843)

and

Yt = 2573.94 + 8.22 It (811.9) (0.843)

=π From example (A.5) in the Appendix, we see that β 12 / π 22 = 7.22/8.22 = 0.878 as described  in (A.24). Also, β = ( π 22 − 1)/ π 22 = (8.22 − 1)/8.22 = 7.22/822 = 0.878 as described in (A.25). Similarly, α =π 11 / π 22 = π 21 / π 22 = 2573.94/8.22 = 313.01 as described in (A.22). This confirms that under just-identification, the 2SLS estimates of the structural coefficients are identical to the Indirect Least Squares (ILS) estimates. The latter estimates uniquely solve for the structural parameter estimates from the reduced form estimates under just-identification. Note that in this case both 2SLS and ILS estimates of the consumption equation are identical  = mci /myi as shown in to the simple IV estimator using It as an instrument for Yt ; i.e., β IV (A.24). 11.2.1

Spatial Lag Dependence

An alternative popular model for spatial lag dependence considered in Section 9.9 is given by: y = ρW y + Xβ + ǫ where ǫ ∼ IIN(0, σ 2 ), see Anselin (1988). Here yi may denote output in region i which is affected by output of its neighbors through the spatial coefficient ρ and the weight matrix W. Recall from section 9.9, W is a known weight matrix with zero elements along its diagonal. It could be a contiguity matrix having elements 1 if its a neighboring region and zero otherwise. Usually this is normalized such that each row sums to 1. Alternatively, W could be based on distances


11.3

System Estimation: Three-Stage Least Squares

267

from neighbors again normalized such that each row sums to 1. It is clear that the presence of W y as a regressor introduces endogeneity. Assuming (In − ρW ) nonsingular, one can solve for the reduced form model: y = (In − ρW )−1 Xβ + ǫ∗ where ǫ∗ = (In − ρW )−1 ǫ has mean zero and variance covariance matrix which has the same form as (9.38), i.e., Σ = E(ǫ∗ ǫ∗′ ) = σ 2 Ω = σ 2 (In − ρW )−1 (In − ρW ′ )−1 For |ρ| < 1, one obtains (In − ρW )−1 = In + ρW + ρ2 W 2 + ρ3 W 3 + ... Hence E(y/X) = (In − ρW )−1 Xβ = Xβ + ρW Xβ + ρ2 W 2 Xβ + ρ3 W 3 Xβ + ... This also means that E(W y/X) = W (In − ρW )−1 Xβ = W Xβ + ρW 2 Xβ + ρ2 W 3 Xβ + ρ3 W 4 Xβ + ... Based on this last expression, Kelejian and Robinson (1993) and Kelejian and Prucha (1998) suggest the use of a subset of the following instrumental variables: {X, W X, W 2 X, W 3 X, W 4 X, ...} Lee (2003) suggested using the optimal instrument matrix:  {X, W (In −  ρW )−1 X β}

 are obtained from a first stage IV estimator, using {X, W X} where the values for  ρ and β as instruments, possibly augmented with W 2 X. Note that Lee’s (2003) instruments involve inverting a matrix of dimension n. Kelejian, et al. (2004) suggest an approximation based upon: {X,

r 

s=0

  ρs W s+1 X β}

where r, the highest order of this approximation depends upon the sample size, with r = o(n1/2 ). In their Monte Carlo experiments, they set r = nc where c = 0.25, 0.35, and 0.45. This is a natural application of 2SLS to deal with the problem of spatial lag dependence.

11.3

System Estimation: Three-Stage Least Squares

If the entire simultaneous equations model is to be estimated, then one should consider system estimators rather than single equation estimators. System estimators take into account the zero restrictions in every equation as well as the variance-covariance matrix of the disturbances of


268

Chapter 11: Simultaneous Equations Model

the whole system. One such system estimator is Three-Stage Least Squares (3SLS) where the structural equations are stacked on top of each other, just like a set of SUR equations, y = Zδ + u

(11.43)

where ⎡

⎤ ⎡ Z1 0 . . . 0 y1 ⎢ y2 ⎥ ⎢ 0 Z2 . . . 0 ⎥ ⎢ y=⎢ ⎣ : ⎦; Z = ⎣ : ... : ... yG 0 ... ZG

⎤ ⎡ δ1 ⎥ ⎢ ⎥ ⎢ ⎥ ; δ = ⎢ δ2 ⎥ ; u = ⎢ ⎦ ⎣ : ⎦ ⎣ δG ⎡

⎤ u1 u2 ⎥ ⎥ : ⎦ uG

and u has zero mean and variance-covariance matrix Σ ⊗ IT , indicating the possible correlation among the disturbances of the different structural equations. Σ = [σ ij ], with E(ui u′j ) = σ ij IT , for i, j = 1, 2, . . . , G. This ⊗ notation was used in Chapter 9 and defined in the Appendix to Chapter 7. Problem 4 shows that premultiplying the i-th structural equation by X ′ and performing GLS on the transformed equation results in 2SLS. For the system given in (11.43), the analogy is obtained by premultiplying by (IG ⊗X ′ ), i.e., each equation by X ′ , and performing GLS on the whole system. The transformed error (IG ⊗ X ′ )u has a zero mean and variancecovariance matrix Σ ⊗ (X ′ X). Hence, GLS on the entire system obtains  δ GLS

= {Z ′ (IG ⊗ X)[Σ−1 ⊗ (X ′ X)−1 ](IG ⊗ X ′ )Z}−1 {Z ′ (IG ⊗ X)[Σ−1 ⊗ (X ′ X)−1 ](IG ⊗ X ′ )y

(11.44)

which upon simplifying yields  δ GLS = {Z ′ [Σ−1 ⊗ PX ]Z}−1 {Z ′ [Σ−1 ⊗ PX ]y}

(11.45)

Σ has to be estimated to make this estimator operational. Zellner and Theil (1962), suggest getting the 2SLS residuals for the i-th equation, say u i = yi − Zi δ i,2SLS and estimating Σ by  Σ = [ σ ij ] where σ ij = [ u′i u j /(T − gi − ki )1/2 (T − gj − kj )1/2 ] for

i, j = 1, 2, . . . , G.

 is substituted for Σ in (11.45), the resulting estimator is called 3SLS: If Σ   −1 ⊗ PX ]Z}−1 {Z ′ [Σ  −1 ⊗ PX ]y} δ 3SLS = {Z ′ [Σ

(11.46)

 −1 ⊗ PX ]Z}−1 . The asymptotic variance-covariance matrix of  δ3SLS can be estimated by {Z ′ [Σ If the system of equations (11.43) is properly specified, 3SLS is more efficient than 2SLS. But if say, the second equation is improperly specified while the first equation is properly specified, then a system estimator like 3SLS will be contaminated by this misspecification whereas a single equation estimator like 2SLS on the first equation is not. So, if the first equation is of interest it does not pay to go to a system estimator in this case. Two sufficient conditions exist for the equivalence of 2SLS and 3SLS, these are the following: (i) Σ is diagonal, and (ii) every equation is just identified. Problem 5 leads you step by step through these results. It is also easy to show, see problem 5, that a necessary and sufficient condition for 3SLS to be equivalent to 2SLS on each equation is given by σ ij Zi′ P¯Zj = 0 for

i, j = 1, 2, . . . , G


11.4

Test for Over-Identification Restrictions

269

where Zi = PX Zi , see Baltagi (1989). This is similar to the condition derived in the seemingly unrelated regressions case except it involves the set of second stage regressors of 2SLS. One can easily see that besides the two sufficient conditions given above, Zi′ P¯Zj = 0 states that the set of second stage regressors of the i-th equation have to be a perfect linear combination of those in the j-th equation and vice versa. A similar condition was derived by Kapteyn and Fiebig (1981). If some equations in the system are over-identified while others are just-identified, the 3SLS estimates of the over-identified equations can be obtained by running 3SLS ignoring the just-identified equations. The 3SLS estimates of each just-identified equation differ from those of 2SLS by a vector which is a linear function of the 3SLS residuals of the over-identified equations, see Theil (1971) and problem 17.

11.4

Test for Over-Identification Restrictions

We emphasized instrument relevance, now we turn to instrument exogeneity. Under justidentification, one cannot statistically test instruments for exogeneity. This choice of exogenous instruments requires making an expert judgement based on knowledge of the empirical application. However, if the first structural equation is over-identified, i.e., the number of instruments ℓ is larger than the number of right hand side variables (g1 + k1 ), then one can test these overidentifying restrictions. A likelihood ratio test for this over-identification condition based on maximum likelihood procedures was given by Anderson and Rubin (1950). This version of the test requires the computation of LIML. This was later modified by Basmann (1960) so that it could be based on the 2SLS procedure. Here we present a simpler alternative based on Davidson and MacKinnon (1993) and Hausman (1983). In essence, one is testing Ho ; y1 = Z1 δ 1 + u1

versus

H1 ; y1 = Z1 δ 1 + W ∗ γ + u1

(11.47)

where u1 ∼ IID(0, σ 11 IT ). Let W be the matrix of instruments of full rank ℓ. Also, let W ∗ be a 1 = PW Z1 . subset of instruments W , of dimension (ℓ−k1 −g1 ), that are linearly independent of Z ∗  In this case, the matrix [Z1 , W ] has full rank ℓ and therefore, spans the same space as W . A test for over-identification is a test for γ = 0. In other words, W ∗ has no ability to explain any variation in y1 that is not explained by Z1 using the matrix of instruments W . If W ∗ is correlated with u1 or the first structural equation (11.34) is misspecified, say by Z1 not including some variables in W ∗ , then γ = 0. Hence, testing γ = 0 should be interpreted as a joint test for the validity of the matrix of instruments W and the proper specification of (11.34) see, Davidson and MacKinnon (1993). Testing Ho ; γ = 0 can be obtained as an asymptotic F -test as follows: (RRSS ∗ − URSS ∗ )/ℓ − (g1 + k1 ) URSS /(T − ℓ)

(11.48)

This is asymptotically distributed as F (ℓ − (g1 + k1 ), T − ℓ) under Ho . Using instruments W, we regress Z1 on W and get Z1 , then obtain the restricted 2SLS estimate  δ 1,2SLS by regressing  y1 on Z1 . The restricted residual sum of squares from the second stage regression is RRSS ∗ = 1 (y1 − Z1 δ1,2SLS )′ (y1 − Z δ 1,2SLS ). Next we regress y1 on Z1 and W ∗ to get the unrestricted 2SLS estimates  δ 1,2SLS and γ 2SLS . The unrestricted residual sum of squares from the second stage ∗ regression is URSS = (y1 − Z1 δ1,2SLS − W ∗  γ 2SLS )′ (y1 − Z1 δ 1,2SLS − W ∗ γ 2SLS ). The URSS


270

Chapter 11: Simultaneous Equations Model

in (11.48) is the 2SLS residuals sum of squares from the unrestricted model which is obtained as follows (y1 − Z1 δ 1,2SLS − W ∗ γ 2SLS )′ (y1 − Z1 δ1,2SLS − W ∗ γ 2SLS ). URSS differs from URSS ∗ in that Z1 rather than Z1 is used in obtaining the residuals. Note that this differs from the Chow-test in that the denominator is not based on URSS ∗ , see Wooldridge (1990). This test does not require the construction of W ∗ for its implementation. This is because the model under H1 is just-identified with as many regressor as there are instruments. This means that its URSS ∗ = y1′ P¯W y1 = y1′ y1 − y1′ PW y1 see problem 10. It is easy to show, see problem 12, that RRSS ∗ = y1′ P¯Z1 y1 = y1′ y1 − y1′ PZ1 y1

1 = PW Z1 . Hence, where Z

RRSS ∗ − URSS ∗ = y1′ PW y1 − y1′ PZ1 y1

(11.49)

δ 1,2SLS )/T δ 1,2SLS )′ (y1 − Z1 σ 11 = (y1 − Z1

(11.50)

The test for over-identification can therefore be based on RRSS ∗ − U RSS ∗ divided by a consistent estimate of σ 11 , say,

Problem 12 shows that the resulting test statistic is exactly that proposed by Hausman (1983). δ1,2SLS In a nutshell, the Hausman over-identification test regresses the 2SLS residuals y1 − Z1 on the matrix W of all pre-determined variables in the model. The test statistic is T times the uncentered R2 of this regression. See the Appendix to Chapter 3 for a definition of uncentered R2 . This test statistic is asymptotically distributed as χ2 with ℓ − (g1 + k1 ) degrees of freedom. Large values of this statistic reject the null hypothesis. Alternatively, one can get this test statistic as a Gauss-Newton Regression (GNR) on the unrestricted model in (11.47). To see this, recall from section 8.4 that the GNR applies to a general nonlinear model yt = xt (β) + ut . Using the set of instruments W , the GNR becomes  = PW X(β)b  + residuals y − x(β)

 denotes the restricted instrumental variable estimate of β under the null hypothesis and where β X(β) is the matrix of derivatives with typical elements Xij (β) = ∂xi (β)/∂β j for j = 1, . . . , k. Thus, the only difference between this GNR and that in Chapter 8 is that the regressors are multiplied by PW , see Davidson and MacKinnon (1993, p. 226). Therefore, the GNR for (11.47) yields 1 b1 + W ∗ b2 + residuals y1 − Z1 δ 1,2SLS = Z

(11.51)

1 , W ∗ ] and  since PW [Z1 , W ∗ ] = [Z δ 1,2SLS is the restricted estimator under Ho ; γ = 0. But, ∗ [Z1 , W ] spans the same space as W , see problem 12. Hence, the GNR in (11.51) is equivalent to running the 2SLS residuals on W and computing T times the uncentered R2 as described above. Once again, it is clear that W ∗ need not be constructed. The basic intuition behind the test for over-identification restriction rests on the fact that one can compute several legitimate IV estimators if all these instruments are relevant and


11.5

Hausman’s Specification Test

271

exogenous. For example, suppose there are two instruments and one right hand side endogenous variable. Then one can compute two IV estimators using each instrument separately. If these IV estimators produce very different estimates, then may be one instrument or the other or both are not exogenous. The over-identification test we just described implicitly makes this comparison without actually computing all possible IV estimates. Exogenous instruments have to be uncorrelated with the disturbances. This suggests that the 2SLS residuals have to be uncorrelated with the instruments. This is the basis for the T Ru2 test statistic. If all the instruments are exogenous, the regression coefficient estimates should all be not significantly different from zero and the Ru2 should be low.

11.5

Hausman’s Specification Test3

A critical assumption for the linear regression model y = Xβ + u is that the set of regressors X are uncorrelated with the error term u. Otherwise, we have simultaneous bias and OLS is inconsistent. Hausman (1978) proposed a general specification test for Ho ; E(u/X) = 0 versus H1 ; E(u/X) = 0. Two estimators are needed to implement this test. The first estimator must be a consistent and efficient estimator of β under Ho which becomes inconsistent under H1 . Let  . The second estimator, denoted by β  , must us denote this efficient estimator under Ho by β o 1 be consistent for β under both Ho and H1 , but inefficient under Ho . Hausman’s test is based  −β  . Under Ho ; plim q is zero, while on the difference between these two estimators q = β 1 o 1 )− var(β o ) and Hausman’s under H1 ; plim q = 0. Hausman (1978) shows that var( q ) = var(β test becomes q )]−1 q m = q′ [var(

(11.52)

which is asymptotically distributed under Ho as χ2k where k is the dimension of β. It remains to show that var( q ) is the difference between the two variances. This can be illustrated for a single regressor case without matrix algebra, see Maddala (1992, page 507).  , q) = 0. To prove this, consider a new estimator of β defined as First, one shows that cov(β o   =β  + λ  = β for every λ and β q where λ is an arbitrary constant. Under Ho , plim β o

 o , q)  = var(β  o ) + λ2 var( var(β) q ) + 2λcov(β

  is efficient, var(β)  ≥ var(β  ) which means that λ2 var(  , q) ≥ 0 for every λ. Since β q )+2λ cov(β o o o  o , q) > 0, then for λ = −cov(β  o , q)/var( If cov(β q ) the above inequality is violated. Similarly, if  o , q)/var(  o , q) < 0, then for λ=-cov(β q ) the above inequality is violated. Therefore, under cov(β  , q) = 0. Ho , for the above inequality to be satisfied for every λ, it must be the case that cov(β o        ). Now, q = β 1 − β o can be rewritten as β 1 = q+ β o with var(β 1 ) = var( q )+ var(β o )+2cov( q, β o   ). Using the fact that the last term is zero, we get the required result: var( q ) = var(β 1 )− var(β o Example 4: Consider the simple regression without a constant yt = βxt + ut

t = 1, 2, . . . , T

with

ut ∼ IIN(0, σ 2 )

and where β is a scalar. Under Ho ; E(ut /xt ) = 0 and OLS is efficient and consistent. Under H1 ; OLS is not consistent. Using wt as an instrumental variable, with wt uncorrelated with ut


272

Chapter 11: Simultaneous Equations Model

and preferably highly correlated with xt , yields the following IV estimator of β:  = T yt wt / T xt wt = β + T wt ut / T xt wt β IV t=1 t=1 t=1 t=1 2  = β under Ho and H1 and var(β  ) = Ď&#x192; T w2 /(T xt wt )2 . Also, with plim β IV IV t=1 t t=1     T T T T 2 2  β OLS = t=1 xt yt / t=1 xt = β + t=1 xt ut / t=1 xt T 2 2    with plim β OLS = β under Ho but plim β OLS under H1 , with var(β OLS ) = Ď&#x192; / t=1 xt . In this IV â&#x2C6;&#x2019; β OLS and plim q = 0 under H1 , while plim q = 0 under Ho , with case, q = β â&#x17D;¤ â&#x17D;Ą T 2 1 â&#x17D;Ľ 2â&#x17D;˘ t=1 wt  ) â&#x2C6;&#x2019; var(β  var( q ) = var(β â&#x17D;Ś 2 â&#x2C6;&#x2019; T IV OLS ) = Ď&#x192; â&#x17D;Ł  2 T t=1 xt t=1 xt

2 1 â&#x2C6;&#x2019; rxw 1 Ď&#x192;2  â&#x2C6;&#x2019; 1 = var(β OLS ) = T 2 2 r2 rxw xw t=1 xt  T T 2 = ( T x w )2 / 2 2 where rxw t=1 t t t=1 wt t=1 xt . Therefore, Hausmanâ&#x20AC;&#x2122;s test statistic is 2 2  /var(β m = q2 rxw OLS )(1 â&#x2C6;&#x2019; rxw )

which is asymptotically distributed as Ď&#x2021;21 under Ho . Note that the same estimator of Ď&#x192; 2 is used 2  ) and var(β  for var(β IV OLS ). This is the estimator of Ď&#x192; obtained under Ho . The Hausman-test can also be obtained from the following augmented regression: xt + ÇŤ t yt = βxt + Îł

where x t is the predicted value of xt from regressing it on the instrumental variable wt . Problem 13 asks the reader to show that Hausmanâ&#x20AC;&#x2122;s test statistic can be obtained by testing Îł = 0. In matrix form, the Durbin-Wu-Hausman test for the ďŹ rst structural equation is based upon the difference between OLS and IV estimation of (11.34) using the matrix of instruments W . In particular, the vector of contrasts is given by δ 1,OLS = (Z â&#x20AC;˛ PW Z1 )â&#x2C6;&#x2019;1 [Z â&#x20AC;˛ PW y1 â&#x2C6;&#x2019; (Z â&#x20AC;˛ PW Z1 )(Z â&#x20AC;˛ Z1 )â&#x2C6;&#x2019;1 Z â&#x20AC;˛ y1 ] (11.53) q =  δ 1,IV â&#x2C6;&#x2019;  1

1

1

1

1

= (Z1â&#x20AC;˛ PW Z1 )â&#x2C6;&#x2019;1 [Z1â&#x20AC;˛ PW PÂŻZ1 y1 ]

Under the null hypothesis, q = (Z1â&#x20AC;˛ PW Z1 )â&#x2C6;&#x2019;1 Z1â&#x20AC;˛ PW PÂŻZ1 u1 . The test for q = 0 can be based on the test for Z1â&#x20AC;˛ PW PÂŻZ1 u1 having mean zero asymptotically. This last vector is of dimension (g1 + k1 ). However, not all of its elements are necessarily random variables since PÂŻZ1 may annihilate some columns of the second stage regressors Z1 = PW Z1 . In fact, all the included Xâ&#x20AC;&#x2122;s which are part of W , i.e., X1 , will be annihilated by PÂŻZ1 . Only the g1 linearly independent variables YË&#x2020;1 = PW Y1 are not annihilated by PÂŻZ1 . â&#x20AC;˛ Our test focuses on the vector YË&#x2020;1 PÂŻZ1 u1 having mean zero asymptotically. Now consider the artiďŹ cial regression y1 = Z1 δ 1 + YË&#x2020;1 Îł + residuals

(11.54)

Since [Z1 , YË&#x2020;1 ], [Z1 , Z1 ], [Z1 , Z1 â&#x2C6;&#x2019; Z1 ] and [Z1 , Y1 â&#x2C6;&#x2019; Y1 ] all span the same column space, this regression has the same sum of squares residuals as y1 = Z1 δ 1 + (Y1 â&#x2C6;&#x2019; YË&#x2020;1 )Ρ + residuals

(11.55)


11.6

Empirical Example: Crime in North Carolina

273

The DWH-test may be based on either of these regressions. It is equivalent to testing γ = 0 in (11.54) or η = 0 in (11.55) using an F -test. This is asymptotically distributed as F (g1 , T − 2g1 − k1 ). Davidson and MacKinnon (1993, p. 239) warn about interpreting this test as one of exogeneity of Y1 (the variables in Z1 not in the space spanned by W ). They argue that what is being tested is the consistency of the OLS estimates of δ1 , not that every column of Z1 is independent of u1 . In practice, one may be sure about using W2 as a set of IV’s but is not sure whether some r additional variables in Z1 are legitimate as instruments. The DWH-test in this case will be δ 2,IV based on W2 and based upon the difference between two IV estimators for δ 1 . The first is  the second is  δ 1,IV based on W1 . The latter set includes W2 and the additional r variables in Z1 .  δ 1,IV δ 2,IV − 

= (Z1′ PW2 Z1 )−1 [Z1′ PW2 y1 − (Z1′ PW2 Z1 )(Z1′ PW1 Z1 )−1 Z1′ PW1 y1 ] )y1 = (Z ′ PW Z1 )−1 Z ′ PW (P¯P 1

2

1

2

(11.56)

W1 Z1

since PW2 PW1 = PW2 . The DWH-test is based on this contrast having mean zero asymptotically. Once again this last vector has dimension g1 + k1 and not all its elements are necessarily random variables since P¯PW1 Z1 annihilates some columns of PW2 Z1 . This test can be based on the following artificial regression: y1 = Z1 δ 1 + PW2 Z1∗ γ + residuals

(11.57)

where PW2 Z1∗ consists of the r columns of PW2 Z1 that are not annihilated by P¯W1 . Regression (11.57) is performed with W1 as the set of IV’s and γ = 0 is tested using an F -test.

11.6

Empirical Example: Crime in North Carolina

Cornwell and Trumbull (1994) estimated an economic model of crime using data on 90 counties in North Carolina observed over the years 1981-87. This data set is available on the Springer web site as CRIME.DAT. Here, we consider cross-section data for 1987 and reconsider the full panel data set in Chapter 12. Table 11.2 gives the OLS estimates relating the crime rate (which is an FBI index measuring the number of crimes divided by the county population) to a set of explanatory variables. This was done using STATA. All variables are in logs except for the regional dummies. The explanatory variables consist of the probability of arrest (which is measured by ratio of arrests to offenses), probability of conviction given arrest (which is measured by the ratio of convictions to arrests), probability of a prison sentence given a conviction (measured by the proportion of total convictions resulting in prison sentences); average prison sentence in days as a proxy for sanction severity. The number of police per capita as a measure of the county’s ability to detect crime, the population density which is the county population divided by county land area, a dummy variable indicating whether the county is in the SMSA with population larger than 50,000. Percent minority, which is the proportion of the county’s population that is minority or non-white. Percent young male which is the proportion of the county’s population that is males and between the ages of 15 and 24. Regional dummies for western and central counties. Opportunities in the legal sector captured by the average weekly wage in the county by industry. These industries are: construction; transportation, utilities and communication; wholesale and retail trade; finance, insurance and real estate; services; manufacturing; and federal, state and local government.


274

Chapter 11: Simultaneous Equations Model

Results show that the probability of arrest as well as conviction given arrest have a negative and significant effect on the crime rate with estimated elasticities of −0.45 and −0.30 respectively. The probability of imprisonment given conviction as well as the sentence severity have a negative but insignificant effect on the crime rate. The greater the number of police per capita, the greater the number of reported crimes per capita. The estimated elasticity is 0.36 and it is significant. This could be explained by the fact that the larger the police force, the larger the reported crime. Alternatively, this could be an endogeneity problem with more crime resulting in the hiring of more police. The higher the density of the population the higher the crime rate. The estimated elasticity is 0.31 and it is significant. Returns to legal activity are insignificant except for wages in the service sector. This has a negative and significant effect on crime with an estimated elasticity of −0.27. Percent young male is insignificant, while percent minority is positive and significant with an estimated elasticity of 0.22. The central dummy variable is negative and significant while the western dummy variable is not significant. Also, the urban dummy variable is insignificant. Cornwell and Trumbull (1994) worried about the endogeneity of police per capita and the probability of arrest. They used as instruments two additional vari-

Table 11.2 Least Squares Estimates: Crime in North Carolina Source

SS

df

MS

Model Residual

22.8072483 3.99245334

20 69

1.14036241 .057861643

Total

26.7997016

89

.301120243

Coef.

Std. Err.

t

P > |t|

−.4522907 −.3003044 −.0340435 −.2134467 .3610463 .3149706 .2727634 .1603777 .1325719 −.3205858 −.2694193 .1029571 .3856593 −.078239 −.1774064 .0326912 .2245975 −.087998 −.1771378 −.0896129 −3.395919

.0816261 .0600259 .1251096 .1167513 .0909534 .0698265 .2198714 .1666014 .3005086 .251185 .1039842 .1524804 .3215442 .2701264 .4251793 .1580377 .0519005 .1243235 .0739535 .1375084 3.020674

−5.54 −5.00 −0.27 −1.83 3.97 4.51 1.24 0.96 0.44 −1.28 −2.59 0.68 1.20 −0.29 −0.42 0.21 4.33 −0.71 −2.40 −0.65 −1.12

0.000 0.000 0.786 0.072 0.000 0.000 0.219 0.339 0.660 0.206 0.012 0.502 0.234 0.773 0.678 0.837 0.000 0.481 0.019 0.517 0.265

lcrmrte lprbarr lprbconv lprbpris lavgsen lpolpc ldensity lwcon lwtuc lwtrd lwfir lwser lwmfg lwfed lwsta lwloc lpctymle lpctmin west central urban cons

Number of obs F(20,69) Prob > F R-squared Adj R-squared Root MSE

= = = = = =

90 19.71 0.0000 0.8510 0.8078 .24054

[95% Conf. Interval] −.6151303 −.4200527 −.2836303 −.4463592 .1795993 .1756705 −.165868 −.171983 −.4669263 −.8216861 −.4768622 −.2012331 −.2558039 −.6171264 −1.025616 −.2825855 .1210589 −.3360167 −.3246709 −.3639347 −9.421998

−.2894511 −.180556 .2155433 .0194659 .5424934 .4542707 .7113949 .4927385 .7320702 .1805146 −.0619765 .4071472 1.027123 .4606485 .670803 .3479678 .3281361 .1600207 −.0296046 .184709 2.630159


11.6

Empirical Example: Crime in North Carolina

275

Table 11.3 Instrumental variables (2SLS) regression: Crime in North Carolina Source

SS

df

MS

Model Residual

22.6350465 4.16465515

20 69

1.13175232 .060357321

Total

26.7997016

89

.301120243

Coef.

Std. Err.

t

P > |t|

−.4393081 .5136133 −.2713278 −.0278416 −.280122 .3273521 .3456183 .1773533 .212578 −.3540903 −.2911556 .0642196 .2974661 .0037846 −.4336541 .0095115 .2285766 −.0952899 −.1792662 −.1139416 −1.159015

.2267579 .1976888 .0847024 .1283276 .1387228 .0893292 .2419206 .1718849 .3239984 .2612516 .1122454 .1644108 .3425026 .3102383 .5166733 .1869867 .0543079 .1301449 .0762815 .143354 3.898202

−1.94 2.60 −3.20 −0.22 −2.02 3.66 1.43 1.03 0.66 −1.36 −2.59 0.39 0.87 0.01 −0.84 0.05 4.21 −0.73 −2.35 −0.79 −0.30

0.057 0.011 0.002 0.829 0.047 0.000 0.158 0.306 0.514 0.180 0.012 0.697 0.388 0.990 0.404 0.960 0.000 0.467 0.022 0.429 0.767

lcrmrte lprbarr lpolpc lprbconv lprbpris lavgsen ldensity lwcon lwtuc lwtrd lwfir lwser lwmfg lwfed lwsta lwloc lpctymle lpctmin west central urban cons Instrumented: Instruments:

Number of obs F(20,69) Prob > F R-squared Adj R-squared Root MSE

= = = = = =

90 17.35 0.0000 0.8446 0.7996 .24568

[95% Conf. Interval] −.8916777 .1192349 −.4403044 −.2838482 −.5568663 .1491452 −.137 −.1655477 −.433781 −.8752731 −.5150789 −.263771 −.3858079 −.615124 −1.464389 −.3635166 .1202354 −.3549219 −.3314437 −.3999251 −8.935716

.0130615 .9079918 −.1023512 .2281651 −.0033776 .505559 .8282366 .5202542 .8589371 .1670925 −.0672322 .3922102 .9807402 .6226931 .597081 .3825397 .3369179 .1643422 −.0270888 .1720419 6.617686

lprbarr lpolpc lprbconv lprbpris lavgsen ldensity lwcon lwtuc lwtrd lwfir lwser lwmfg lwfed lwsta lwloc lpctymle lpctmin west central ltaxpc lmix

ables. Offense mix which is the ratio of crimes involving face to face contact (such as robbery, assault and rape) to those that do not. The rationale for using this variable is that arrest is facilitated by positive identification of the offender. The second instrument is per capita tax revenue. This is justified on the basis that counties with preferences for law enforcement will vote for higher taxes to fund a larger police force. The 2SLS estimates are reported in Table 11.3. The probability of arrest has an estimated elasticity of −0.44 but now with a p-value of 0.057. The probability of conviction given arrest has an estimated elasticity of −0.27 still significant. The probability of imprisonment given conviction is still insignificant while the sentence severity is now negative and significant with an estimated elasticity of −0.28. Police per capita has a higher elasticity of 0.51 still significant. The remaining estimates are slightly affected. In fact, the Hausman test based on the difference between the OLS and 2SLS estimates is shown in Table 11.4. This is computed


276

Chapter 11: Simultaneous Equations Model

Table 11.4 Hausman’s Test: Crime in North Carolina Coefficients

lprbarr lpolpc lprbconv lprbpris lavgsen ldensity lwcon lwtuc lwtrd lwfir lwser lwmfg lwfed lwsta lwloc lpctymle lpctmin west central urban

(b) b2sls

(B) bols

(b–B) Difference

sqrt(diag(V b–V B)) S.E.

−.4393081 .5136133 −.2713278 −.0278416 −.280122 .3273521 .3456183 .1773533 .212578 −.3540903 −.2911556 .0642196 .2974661 .0037846 −.4336541 .0095115 .2285766 −.0952899 −.1792662 −.1139416

−.4522907 .3610463 −.3003044 −.0340435 −.2134467 .3149706 .2727634 .1603777 .1325719 −.3205858 −.2694193 .1029571 .3856593 −.078239 −.1774064 .0326912 .2245975 −.087998 −.1771378 −.0896129

.0129826 .152567 .0289765 .0062019 −.0666753 .0123815 .0728548 .0169755 .0800061 −.0335045 −.0217362 −.0387375 −.0881932 .0820236 −.2562477 −.0231796 .0039792 −.0072919 −.0021284 −.0243287

.2115569 .1755231 .0597611 .0285582 .0749208 .0557132 .1009065 .0422893 .1211178 .0718228 .0422646 .0614869 .1179718 .1525764 .293554 .0999404 .0159902 .0384885 .0187016 .0405192

b = consistent under Ho and Ha; obtained from ivreg B = inconsistent under Ha, efficient under Ho; obtained from regress Test: Ho: difference in coefficients not systematic chi2(20) = (b−B)’[(V b−V B)ˆ(−1)](b−B) = 0.87 Prob > chi2 = 1.0000

using STATA and it contrasts 20 slope coefficient estimates. The Hausman test statistic is 0.87 and is asymptotically distributed as χ220 . This is insignificant, and shows that the 2SLS and OLS estimates are not significantly different given this model specification and the specific choice of instruments. Note that this is a just-identified equation and one cannot test for over-identification. Tables 11.5 and 11.6 give the first-stage regressions for police per capita and the probability of arrest. The R2 of these regressions are 0.56 and 0.47, respectively. The F -statistics for the significance of all slope coefficients are 4.42 and 3.11, respectively. The additional instruments (offense mix and per capita tax revenue) are jointly significant in both regressions yielding F statistics of 10.56 and 5.78 with p-values of 0.0001 and 0.0048, respectively. Although there are two right hand side endogenous regressors in the crime equation rather than one, the Stock and Watson ‘rule of thumb’ suggest that these instruments may be weak.


Notes

277

Table 11.5 First Stage Regression: Police per Capita Source

SS

df

MS

Model Residual

6.99830344 5.46683312

20 69

0.349915172 0.079229465

Total

12.4651366

89

0.140057714

Coef.

Std. Err.

t

P > |t|

0.2177256 0.5601989 0.0037744 −0.0487064 0.3958972 0.0201292 −0.5368469 −0.0216638 −0.4207274 0.0001257 0.0973089 0.1710295 0.8555422 −0.1118764 1.375102 0.4186939 −0.0517966 0.1458865 0.0477227 −0.1192027 −16.33148

0.0733414 0.1489398 0.0744622 0.1470085 0.1295763 0.0798454 0.2562641 0.1955598 0.3483584 0.2976009 0.1272819 0.1814396 0.3779595 0.3181352 0.4676561 0.1869473 0.0619159 0.1490133 0.0877814 0.1719407 3.221824

2.97 3.76 0.05 −0.33 3.06 0.25 −2.09 −0.11 −1.21 0.00 0.76 0.94 2.26 −0.35 2.94 2.24 −0.84 0.98 0.54 −0.69 −5.07

0.004 0.000 0.960 0.741 0.003 0.802 0.040 0.912 0.231 1.000 0.447 0.349 0.027 0.726 0.004 0.028 0.406 0.331 0.588 0.490 0.000

lpolpc lmix ltaxpc lprbconv lprbpris lavgsen ldensity lwcon lwtuc lwtrd lwfir lwser lwmfg lwfed lwsta lwloc lpctymle lpctmin west central urban cons

Number of obs F(20,69) Prob > F R-squared Adj R-squared Root MSE

= 90 = 4.42 = 0.0000 = 0.5614 = 0.4343 = 0.28148

[95% Conf. Interval] 0.0714135 0.2630721 −0.1447736 −0.3419802 0.1373996 −0.1391581 −1.04808 −0.411795 −1.115683 −0.5935718 −0.1566116 −0.1909327 0.1015338 −0.7465387 0.4421535 0.0457442 −0.1753154 −0.151387 −0.1273964 −0.4622151 −22.75884

0.3640378 0.8573258 0.1523223 0.2445675 0.6543948 0.1794165 −0.025614 0.3684674 0.2742286 0.5938232 0.3512293 0.5329916 1.609551 0.5227859 2.30805 0.7916436 0.0717222 0.4431599 0.2228419 0.2238097 −9.904113

Notes 1. A heteroskedasticity-robust statistic is recommended especially if y2 has discrete characteristics. 2. Why 10? See the proof in Appendix 10.4 of Stock and Watson (2003). 3. This test is also known as the Durbin-Wu-Hausman test, following the work of Durbin (1954), Wu (1973) and Hausman (1978).

Problems 1. Show that the OLS estimator of δ in (11.14), which can be written as T T  δ OLS = t=1 pt qt / t=1 p2t T T is not consistent for δ. Hint: Write  δ OLS = δ + t=1 pt (u2t − u¯2 )/ t=1 p2t , and use (11.18) to show that plim  δ OLS = δ + (σ 12 − σ 22 )(δ − β)/[σ 11 + σ 22 − 2σ 12 ].


278

Chapter 11: Simultaneous Equations Model

Table 11.6 First Stage Regression: Probability of Arrest Source

SS

df

MS

Model Residual

6.84874028 7.59345096

20 69

0.342437014 0.110050014

Total

14.4421912

89

0.162271812

Coef.

Std. Err.

t

P > |t|

0.2682143 −0.1938134 −0.1946392 −0.0240173 0.1565061 −0.2211654 −0.2024569 −0.0461931 0.0494793 0.050559 0.0551851 0.0550689 0.2622408 −0.4843599 0.7739819 −0.3373594 −0.0096724 0.0701236 0.0112086 −0.0150372 −4.319234

0.0864373 0.1755345 0.0877581 0.1732583 0.1527134 0.0941026 0.3020226 0.230479 0.4105612 0.3507405 0.1500094 0.2138375 0.4454479 0.3749414 0.5511607 0.2203286 0.0729716 0.1756211 0.1034557 0.2026425 3.797113

3.10 −1.10 −2.22 −0.14 1.02 −2.35 −0.67 −0.20 0.12 0.14 0.37 0.26 0.59 −1.29 1.40 −1.53 −0.13 0.40 0.11 −0.07 −1.14

0.003 0.273 0.030 0.890 0.309 0.022 0.505 0.842 0.904 0.886 0.714 0.798 0.558 0.201 0.165 0.130 0.895 0.691 0.914 0.941 0.259

lcrmrte lmix ltaxpc lprbconv lprbpris lavgsen ldensity lwcon lwtuc lwtrd lwfir lwser lwmfg lwfed lwsta lwloc lpctymle lpctmin west central urban cons

Number of obs F(20,69) Prob > F R-squared Adj R-squared Root MSE

= 90 = 3.11 = 0.0002 = 0.4742 = 0.3218 = 0.33174

[95% Conf. Interval] 0.0957766 −0.5439952 −0.3697119 −0.3696581 −0.1481488 −0.408895 −0.8049755 −0.5059861 −0.769568 −0.6491491 −0.2440754 −0.3715252 −0.6264034 −1.232347 −0.3255536 −0.776903 −0.1552467 −0.280231 −0.1951798 −0.4192979 −11.89427

0.4406519 0.1563684 −0.0195665 0.3216236 0.4611611 −0.0334357 0.4000616 0.4135999 0.8685266 0.7502671 0.3544456 0.481663 1.150885 0.2636277 1.873517 0.1021842 0.1359019 0.4204782 0.217597 0.3892234 3.255798

2. Consider equation (11.30) and let X3 and X4 be the only two other exogenous variables in this system. (a) Show that a two-stage estimator which regresses y2 on X1 , X2 and X3 to get y2 = yˆ2 + vˆ2 , and y3 on X1 , X2 and X4 to get y3 = yˆ3 + vˆ3 , and then regresses y1 on yˆ2 , yˆ3 and X1 and estimators. Hint: X2 does not necessarily yield consistent TShow that the composite error is T ǫ1 = (u1 + α12 vˆ2 + α13 vˆ3 ) and t=1 ˆǫ1t yˆ2t = 0, because t=1 yˆ2t vˆ3t = 0. The latter does not  hold because Tt=1 X3t vˆ3t = 0. (This shows that if both y’s are not regressed on the same set of X’s, the resulting two stage regression estimates are not consistent).

(b) Show that the two-stage estimator which regresses y2 and y3 on X2 , X3 and X4 to get y2 = yˆ2 + vˆ2 and y3 = yˆ3 + vˆ3 and then regresses y1 on yˆ2 , yˆ3 and X1 and X2 is not necessarily consistent. the composite error term T ǫ1 = u1 + α12 vˆ2 + α13 vˆ3 T Hint: Show that T does not satisfy t=1 ˆǫ1t X1t = 0, since t=1 vˆ2t X1t = 0 and t=1 vˆ3t X1t = 0. (This shows that if one of the included X’s is not included in the first stage regression, then the resulting two-stage regression estimates are not consistent). 3. If equation (11.34) is just-identified, then X2 is of the same dimension as Y1 , i.e., both are T × g1 . Hence, Z1 is of the same dimension as X, both of dimension T × (g1 + k1 ). Therefore, X ′ Z1 is a square nonsingular matrix of dimension (g1 + k1 ). Hence, (Z1′ X)−1 exists. Using this fact, show


Problems

279

that  δ 1,2SLS given by (11.36) reduces to (X ′ Z1 )−1 X ′ y1 . This is exactly the IV estimator with W = X, given in (11.41). Note that this is only feasible if X ′ Z1 is square and nonsingular.

4. Premultiply equation (11.34) by X ′ and show that the transformed disturbances X ′ u1 ∼ (0, σ 11 (X ′ X)). Perform GLS on the resulting transformed equation and show that  δ 1,GLS is  δ 1,2SLS , given by (11.36). 5. The Equivalence of 3SLS and 2SLS.

δ 2SLS , when (i) Σ is diagonal, or (ii) every equa(a) Show that  δ 3SLS given in (11.46) reduces to   −1 ⊗ PX is block-diagonal with tion in the system is just-identified. Hint: For (i); show that Σ  −1 ⊗ PX ]Z}−1 σ ii . Also, Z is block-diagonal, therefore, {Z ′ [Σ the i-th block consisting of PX / is block-diagonal with the i-th block consisting of σ ii (Zi′ PX Zi )−1 . Similarly, computing  −1 ⊗ PX ]y, one can show that the i-th element of  Z ′ [Σ δ 3SLS is (Zi′ PX Zi )−1 Zi′ PX yi = ′  δ i,2SLS . For (ii); show that Zi X is square and nonsingular under just-identification. Therefore,  δ i,2SLS = (X ′ Zi )−1 X ′ yi from problem 3. Also, from (11.44), we get  δ 3SLS

=

 −1 ⊗ (X ′ X)−1 )diag[X ′ Zi ]}−1 {diag[Zi′ X](Σ ′  −1 ⊗ (X ′ X)−1 )(IG ⊗ X ′ )y}. {diag[Z X](Σ

Using the fact that

i

Zi′ X

is square, one can show that  δ i,3SLS = (X ′ Zi )−1 X ′ yi .

(b) Premultiply the system of equations in (11.43) by (IG ⊗ PX ) and let y ∗ = (IG ⊗ PX )y, Z ∗ = (IG ⊗ PX )Z and u∗ = (IG ⊗ PX )u, then y ∗ = Z ∗ δ + u∗ .Show that OLS on this transformed model yields 2SLS on each equation in (11.43). Show that GLS on this model yields 3SLS (knowing the true Σ) given in (11.45). Note that var(u∗ ) = Σ ⊗ PX and its generalized inverse is Σ−1 ⊗ PX . Use the Milliken and Albohali condition for the equivalence of OLS and GLS given in equation (9.7) of Chapter 9 to deduce that 3SLS is equivalent to 2SLS if Z ∗′ (Σ−1 ⊗ PX )P¯Z ∗ = 0. Show that this reduces to the following necessary and ′ P¯  = 0 for i = j, see Baltagi (1989). Hint: Use the fact that sufficient condition σ ij Z i Zj i ] and P¯Z ∗ = diag[P¯  ]. Z ∗ = diag[PX Zi ] = diag[Z Zi

Verify that the two sufficient conditions given in part (a) satisfy this necessary and sufficient condition. 6. Consider the following demand and supply equations: Q = Q =

a − bP + u1 c + dP + eW + f L + u2

where W denotes weather conditions affecting supply, and L denotes the supply of immigrant workers available at harvest time. (a) Write this system in the matrix form given by equation (A.1) in the Appendix. (b) What does the order-condition for identification say about these two equations? (c) Premultiply this system by a nonsingular matrix F = [fij ], for i, j = 1, 2. What restrictions must the matrix F satisfy if the transformed model is to satisfy the same restrictions of the original model? Show that the first row of F is in fact the first row of an identity matrix, but the second row of F is not the second row of an identity matrix. What do you conclude?


280

Chapter 11: Simultaneous Equations Model

7. Answer the same questions in problem 6 for the following model: Q = Q =

a − bP + cY + dA + u1 e + f P + gW + hL + u2

where Y is real income and A is real assets. 8. Consider example (A.1) in the Appendix. Recall, that system of equations (A.3) and (A.4) are just-identified. (a) Construct φ for the demand equation (A.3) and show that Aφ = (0, −f )′ which is of rank 1 as long as f = 0. Similarly, construct φ for the supply equation (A.4) and show that Aφ = (−c, 0)′ which is of rank 1 as long as c = 0.

(b) Using equation (A.17), show how the structural parameters can be retrieved from the reduced form parameters. Derive the reduced form equations for this system and verify the above relationships relating the reduced form and structural form parameters.

9. Derive the reduced form equations for the model given in problem 6, and show that the structural parameters of the second equation cannot be derived from the reduced form parameters. Also, show that there are more than one way of expressing the structural parameters of the first equation in terms of the reduced form parameters. 10. Just-Identified Model. Consider the just-identified equation y1 = Z1 δ 1 + u 1 with W , the matrix of instruments for this equation of dimension T × ℓ where ℓ = g1 + k1 the dimension of Z1 . In this case, W ′ Z1 is square and nonsingular. (a) Show that the generalized instrumental variable estimator given below (11.41) reduces to the simple instrumental variable estimator given in (11.38). (b) Show that the minimized value of the criterion function for this just-identified model is zero, i.e., show that (y1 − Z1 δ 1,IV )′ PW (y1 − Z1 δ 1,IV ) = 0. (c) Conclude that the residual sum of squares of the second stage regression of this just-identified model is the same as that obtained by regressing y1 on the matrix of instruments W , i.e., 1 1 = PW Z1 . Hint: Show that show that (y1 − Z δ 1,IV )′ (y1 − Z1 δ 1,IV ) = y1′ P¯W y1 where Z PZ1 = PPW Z1 = PW , under just-identification.

11. Let W1 and W2 be two sets of instrumental variables for the first structural equation given in (11.34). Suppose that W1 is spanned by the space of W2 . Verify that the resulting IV estimator of δ 1 based on W2 is at least as efficient as that based on W1 . Hint: Show that PW2 W1 = W1 and that PW2 − PW1 is idempotent. Conclude that the difference in the corresponding asymptotic covariances of these IV estimators is positive semi-definite. (This shows that increasing the number of legitimate instruments should improve the asymptotic efficiency of an IV estimator). 12. Testing for Over-Identification. In testing Ho ; γ = 0 versus H1 ; γ = 0 in section 11.4, equation (11.47): (a) Show that the second stage regression of 2SLS on the unrestricted model y1 = Z1 δ 1 +W ∗ γ+u1 with the matrix of instruments W yields the following residual sum of squares: URSS ∗ = y1′ P¯W y1 = y1′ y1 − y1′ PW y1 Hint: Use the results of problem 10 for the just-identified case.


Problems

281

(b) Show that the second stage regression of 2SLS on the restricted model y1 = Z1 δ 1 + u1 with the matrix of instruments W yields the following residual sum of squares: RRSS â&#x2C6;&#x2014; = y1â&#x20AC;˛ PÂŻZ1 y1 = y1â&#x20AC;˛ y1 â&#x2C6;&#x2019; y1â&#x20AC;˛ PZ1 y1 where Z1 = PW Z1 and PZ1 = PW Z1 (Z1â&#x20AC;˛ PW Z1 )â&#x2C6;&#x2019;1 Z1â&#x20AC;˛ PW . Conclude that RRSS â&#x2C6;&#x2014; â&#x2C6;&#x2019; URSS â&#x2C6;&#x2014; yields (11.49).

(c) Consider the test statistic (RRSS â&#x2C6;&#x2014; â&#x2C6;&#x2019; URSS â&#x2C6;&#x2014;)/ Ď&#x192; 11 where Ď&#x192;  11 is given by (11.50) as the usual 2SLS residual sum of squares under Ho divided by T . Show that it can be written as Hausmanâ&#x20AC;&#x2122;s (1983) test statistic, i.e., nRu2 where Ru2 is the uncentered R2 of the regression of 2SLS residuals (y1 â&#x2C6;&#x2019; Z1 δ 1,2SLS ) on the matrix of all pre-determined variables W . Hint: Show that δ 1,2SLS )â&#x20AC;˛ PW (y1 â&#x2C6;&#x2019; Z1 δ 1,2SLS ) = (RRSS â&#x2C6;&#x2014;â&#x2C6;&#x2019; URSS â&#x2C6;&#x2014; ) the regression sum of squares (y1 â&#x2C6;&#x2019; Z1 given in (11.49).

(d) Verify that the test for Ho based on the GNR for the model given in part (a) yields the same T Ru2 test statistic described in part (c).

13. Hausmanâ&#x20AC;&#x2122;s SpeciďŹ cation Test: OLS Versus 2SLS. This is based on Maddala (1992, page 511). For the simple regression yt = βxt + ut

t = 1, 2 . . . , T

where β is scalar and ut â&#x2C6;ź IIN(0, Ď&#x192; 2 ). Let wt be an instrumental variable for xt . Run xt on wt and get xt = Ď&#x20AC; wt +  ν t or xt = x Ë&#x2020;t +  ν t where x Ë&#x2020;t = Ď&#x20AC; wt .

(a) Show that in the augmented regression yt = βxt + Îł x Ë&#x2020;t + ÇŤt a test for Îł = 0 based on OLS 2 ) from this regression yields Hausmanâ&#x20AC;&#x2122;s test-statistic. Hint: Show that Îł OLS = q/(1 â&#x2C6;&#x2019; rxw where 2   T T T 2 = / t=1 wt2 t=1 x2t . rxw t=1 xt wt 2 2  Next, show that var( Îł OLS ) = var(β OLS )/rxw (1 â&#x2C6;&#x2019; rxw ). Conclude that 2 2  2OLS /var( Îł Îł OLS ) = q2 rxw /[var(β OLS )(1 â&#x2C6;&#x2019; rxw )]

is the Hausman (1978) test statistic m given in section 11.5. (b) Show that the same result in part (a) could have been obtained from the augmented regression yt = βxt + γ ν t + Ρt where  ν t is the residual from the regression of xt on wt .

14. Consider the following structural equation: y1 = Îą12 y2 + Îą13 y3 + β 11 X1 + β 12 X2 + u1 where y2 and y3 are endogenous and X1 and X2 are exogenous. Also, suppose that the excluded exogenous variables include X3 and X4 . (a) Test the null hypothesis that y2 and y3 are exogenous against the alternative that they are not. Show that Hausmanâ&#x20AC;&#x2122;s test statistic can be obtained from the augmented regression: y1 = Îą12 y2 + Îą13 y3 + β 11 X1 + β 12 X2 + Îł 2 yË&#x2020;2 + Îł 3 yË&#x2020;3 + ÇŤ1 where yË&#x2020;2 and yË&#x2020;3 are predicted values from regressing y2 and y3 on X = [X1 ,X2 ,X3 ,X4 ]. Hausmanâ&#x20AC;&#x2122;s test is equivalent to testing Ho ; Îł 2 = Îł 3 = 0. See equation (11.54).


282

Chapter 11: Simultaneous Equations Model (b) Show that the same results in part (a) hold if we had used the following augmented regression: y1 = α12 y2 + α13 y3 + β 11 X1 + β 12 X2 + γ 2  ν2 + γ3 ν 3 + η1

where  ν 2 and  ν 3 are the residuals from running y2 and y3 on X = [X1 , X2 , X3 , X4 ]. See equation (11.55). Hint: Show that the regressions in (a) and (b) have the same residual sum of squares. 15. For the artificial regression given in (11.55): δ 1,IV = (Z1′ PW Z1 )−1 Z1′ PW y1 . Hint: Y1 − Yˆ1 = (a) Show that OLS on this model yields  δ 1,OLS =  ¯ PW Y1 . Use the FWL Theorem to residual out these variables in (11.55) and use the fact that P¯W Z1 = [P¯W Y1 , 0]. (b) Show that the var( δ 1,OLS ) = s11 (Z1′ PW Z1 )−1 where s11 is the mean squared error of the OLS regression in (11.55). Note that when η = 0 in (11.55), IV estimation is necessary and δ 1,IV )′ (y1 − Z1 δ 1,IV )/T . s11 underestimates σ 11 and will have to be replaced by (y1 − Z1

16. Recursive Systems. A recursive system has two crucial features: B is a triangular matrix and Σ is a diagonal matrix. For this special case of the simultaneous equations model, OLS is still consistent, and under normality of the disturbances still maximum likelihood. Let us consider a specific example: y1t + γ 11 x1t + γ 12 x2t β 21 y1t + y2t + γ 23 x3t In this case, B =

1 β 21

0 1

= =

u1t u2t

is triangular and



=

σ 11 0

0 σ 22

is assumed diagonal.

(a) Check the identifiability conditions of this recursive system. (b) Solve for the reduced form and show that y1t is only a function of the xt ’s and u1t , while y2t is a function of the xt ’s and a linear combination of u1t and u2t . (c) Show that OLS on the first structural equation yields consistent estimates. Hint: There are no right hand side y’s for the first equation. Show that despite the presence of y1 in the second equation, OLS of y2 on y1 and x3 yields consistent estimates. Note that y1 is a function of u1 only and u1 and u2 are not correlated. (d) Under the normality assumption on the disturbances, the likelihood function conditional on the x’s is given by L(B, Γ, Σ) = (2π)−T /2 |B|T |Σ|−T /2 exp(−

1 T u′ Σ−1 ut ) 2 t=1 t

where in this two equation case u′t = (u1t , u2t ). Since B is triangular, |B| = 1. Show that T ′ −1 ut . maximizing L with respect to B and Γ is equivalent to minimizing Q = t=1 ut Σ   T T Conclude that when Σ is diagonal, Σ−1 is diagonal and Q = t=1 u21t /σ 11 + t=1 u22t /σ 22 . Hence, maximizing the likelihood with respect to B and Γ is equivalent to running OLS on each equation separately. 17. Hausman’s Specification Test: 2SLS Versus 3SLS. This is based on Holly (1988). Consider the two-equations model, y1 y2

= αy2 + β 1 x1 + β 2 x2 + u1 = γy1 + β 3 x3 + u2


Problems

283

where y1 and y2 are endogenous; x1 , x2 and x3 are exogenous (the y’s and the x’s are n × 1 vectors). The standard assumptions are made on the disturbance vectors u1 and u2 . With the usual notation, the model can also be written as y1

= Z1 δ 1 + u 1

y2

= Z2 δ 2 + u 2

 The following notation will be used:  δ = 2SLS,  δ = 3SLS, and the corresponding residuals will  be denoted as u  and u , respectively. (a) Assume that αγ = 1. Show that the 3SLS estimating equations reduce to   σ 11 X ′ u 12 X ′ u 1 + σ 2 = 0 12 ′ 22 ′   σ  Z2 PX u  Z2 PX u 1 + σ 2 = 0

where X = (x1 , x2 , x3 ), Σ = [σ ij ] is the structural form covariance matrix, and Σ−1 = [σ ij ] for i, j = 1, 2.   δ2 =  δ1 =  (b) Deduce that  δ 1 − ( σ 12 / σ 22 )(Z1′ PX Z1 )−1 Z1′ PX u 2 . This proves that the δ 2 and  3SLS estimator of the over-identified second equation is equal to its 2SLS counterpart. Also, the 3SLS estimator of the just-identified first equation differs from its 2SLS (or indirect least squares) counterpart by a linear combination of the 2SLS (or 3SLS) residuals of the over-identified equation, see Theil (1971).  δ 1 ? Show that it δ 1 and  (c) How would you interpret a Hausman-type test where you compare  2 2 is nR where R is the R-squared of the regression of u 2 on the set of second stage regressors 2 . Hint: See the solution by Baltagi (1989). 1 and Z of both equations Z

18. For the two-equation simultaneous model

y1t = β 12 y2t + γ 11 x1t + u1t y2t = β 21 y1t + γ 22 x2t + γ 23 x3t + u2t With ⎤ ⎤ ⎡

5 10 20 0 0 3 4 X ′ X = ⎣ 0 20 0 ⎦ X ′ Y = ⎣ 40 20 ⎦ Y ′ Y = 4 8 0 0 10 20 30 ⎡

(a) Determine the identifiability of each equation with the aid of the order and rank conditions for identification. (b) Obtain the OLS normal equations for both equations. Solve for the OLS estimates. (c) Obtain the 2SLS normal equations for both equations. Solve for the 2SLS estimates. (d) Can you estimate these equations using Indirect Least Squares? Explain. 19. Laffer (1970) considered the following supply and demand equations for Traded Money: log(T M/P ) = αo + α1 log(RM/P ) + α2 log i + u1 log(T M/P ) = β o + β 1 log(Y /P ) + β 2 log i + β 3 log(S1) + β 4 log(S2) + u2


284

Chapter 11: Simultaneous Equations Model where TM RM

= Nominal total trade money = Nominal effective reserve money

Y = GNP in current dollars S2 = Degree of market utilization i = short-term rate of interest S1 = Mean real size of the representative economic unit (1939 = 100) P

= GNP price deflator (1958 = 100)

The basic idea is that trade credit is a line of credit and the unused portion represents purchasing power which can be used as a medium of exchange for goods and services. Hence, Laffer (1970) suggests that trade credit should be counted as part of the money supply. Besides real income and the short-term interest rate, the demand for real traded money includes log(S1) and log(S2). S1 is included to capture economies of scale. As S1 increases, holding everything else constant, the presence of economies of scale would mean that the demand for traded money would decrease. Also, the larger S2, the larger the degree of market utilization and the more money is needed for transaction purposes. The data are provided on the Springer web site as LAFFER.ASC. This data covers 21 annual observations over the period 1946-1966. This was obtained from Lott and Ray (1992). Assume that (T M/P ) and i are endogenous and the rest of the variables in this model are exogenous. (a) Using the order condition for identification, determine whether the demand and supply equations are identified? What happens if you used the rank condition of identification? (b) Estimate this model using OLS. (c) Estimate this model using 2SLS. (d) Estimate this model using 3SLS. Compare the estimates and their standard errors for parts (b), (c) and (d). (e) Test the over-identification restriction of each equation. (f) Run Hausman’s specification test on each equation basing it on OLS and 2SLS. (g) Run Hausman’s specification test on each equation basing it on 2SLS and 3SLS. 20. The market for a certain good is expressed by the following equations: Dt St

= α0 − α1 Pt + α2 Xt + u1t (α1 , α2 > 0) = β 0 + β 1 Pt + u2t (β 1 > 0)

Dt

= St = Q t

where Dt is the quantity demanded, St is the quantity supplied, Xt is an exogenous demand shift variable. (u1t , u2t ) is an IID random vector with zero mean and covariance matrix Σ = [σ ij ] for i, j = 1, 2. (a) Examine the identifiability of the model under the assumptions given above using the order and rank conditions of identification. (b) Assuming the moment matrix of exogenous variables converge to a finite non-zero matrix, derive the simultaneous equation bias in the OLS estimator of β 1 . (c) If σ 12 = 0 would you expect this bias to be positive or negative? Explain.


Problems

285

21. Consider the following three equations simultaneous model y1

= α 1 + β 2 y2 + γ 1 X1 + u 1

(1)

y2 y3

= α 2 + β 1 y1 + β 3 y3 + γ 2 X2 + u 2 = α 3 + γ 3 X3 + γ 4 X4 + γ 5 X5 + u 3

(2) (3)

where the X’s are exogenous and the y’s are endogenous. (a) Examine the identifiability of this system using the order and rank conditions. (b) How would you estimate equation (2) by 2SLS? Describe your procedure step by step. (c) Suppose that equation (1) was estimated by running y2 on a constant X2 and X3 and the resulting predicted yˆ2 was substituted in (1), and OLS performed on the resulting model. Would this estimating procedure yield consistent estimates of α1 , β 2 and γ 1 ? Explain your answer. (d) How would you test for the over-identification restrictions in equation (1)? 22. Equivariance of Instrumental Variables Estimators. This is based on Sapra (1997). For the structural equation given in (11.34), let the matrix of instruments W be of dimension T × ℓ where ℓ ≥ g1 + k1 as described below (11.41). Then the corresponding instrumental variable estimator δ 1,IV (y1 ) = (Z1′ PW Z1 )−1 Z1′ PW y1 . of δ 1 given below (11.41) is  (a) Show that this IV estimator is an equivariant estimator of δ 1 , i.e., show that for any linear transformation y1∗ = ay1 + Z1 b where a is a positive scalar and b is an (ℓ × 1) real vector, the following relationship holds:  δ 1,IV (y1 ) + b. δ 1,IV (y1∗ ) = a

(b) Show that the variance estimator σ 2 (y1 ) = (y1 − Z1 δ 1,IV (y1 ))′ (y1 − Z1 δ 1,IV (y1 ))/T

is equivariant for σ 2 , i.e., show that σ  2 (y1∗ ) = a2 σ 2 (y1 ).

23. Identification and Estimation of a Simple Two-Equation Model. This is based on Holly (1987). Consider the following two equation model yt1 yt2

= α + βyt2 + ut1 = γ + yt1 + ut2

where the y’s are endogenous variables and the u’s are serially independent disturbances that are identically distributed with zero means and nonsingular covariance matrix Σ = [σ ij ] where E(uti utj ) = σ ij for i, j = 1, 2, and all t = 1, 2, . . . , T. The reduced form equations are given by yt1 = π 11 + ν t1

and yt2 = π 21 + ν t2

with Ω = [ω ij ] where E(ν ti ν tj ) = ωij for i, j = 1, 2 and all t = 1, 2, . . . , T. (a) Examine the identification of this system of two equations when no further information is available. (b) Repeat part (a) when σ 12 = 0.  (c) Assuming σ 12 = 0, show that β OLS , the OLS estimator of β in the first equation is not consistent.


286

Chapter 11: Simultaneous Equations Model (d) Assuming σ 12 = 0, show that an alternative consistent estimator of β is an IV estimator using zt = [(yt2 − y¯2 ) − (yt1 − y¯1 )] as an instrument for yt2 . (e) Show that the IV estimator of β obtained from part (d) is also an indirect least squares estimator of β. Hint: See the solution by Singh and Bhat (1988).

24. Errors in Measurement and the Wald (1940) Estimator. This is based on Farebrother (1985). Let yi∗ be permanent consumption and X ∗ be permanent income, both are measured with error: yi∗ = βx∗i

yi = yi∗ + ǫi

where

and xi = x∗i + ui

for i = 1, 2, . . . , n.

Let x∗i , ǫi and ui be independent normal random variables with zero means and variances σ 2∗ , σ 2ǫ and σ 2u , respectively. Wald (1940) suggested the following estimator of β: Order the sample by y1 , x ¯1 ) be the sample mean of the first half of the the xi ’s and split the sample into two. Let (¯ sample and (¯ y2 , x ¯2 ) be the sample mean of the second half of this sample. Wald’s estimator of β is  = (¯ β y2 − y¯1 )/(¯ x2 − x ¯1 ). It is the slope of the line joining these two sample mean observations. W  can be interpreted as a simple IV estimator with instrument (a) Show that β W zi

=

1 for

= −1

xi ≥ median(x)

for xi < median(x)

where median(x) is the sample median of x1 , x2 , . . . , xn . (b) Define wi = ρ2 x∗i − τ 2 ui where ρ2 = σ 2u /(σ 2u + σ 2∗ ) and τ 2 = σ 2∗ /(σ 2u + σ 2∗ ). Show that E(xi wi ) = 0 and that wi ∼ N (0, σ 2∗ σ 2u /(σ 2∗ + σ 2u )). (c) Show that x∗i = τ 2 xi + wi and use it to show that 2  /x1 , . . . , xn ) = E(β  E(β W OLS /x1 , . . . , xn ) = βτ .

  Conclude that the exact small sample bias of β OLS and β W are the same.

25. Comparison of t-ratios. This is based on Holly (1990). Consider the two equations model y1 = αy2 + Xβ + u1

and y2 = γy1 + Xβ + u2

where α and γ are scalars, y1 and y2 are T × 1 and X is a T × (K − 1) matrix of exogenous variables. Assume that ui ∼ N (0, σ 2i IT ) for i = 1, 2. Show that the t-ratios for Hoa ; α = 0 and Hob ; OLS are the same. Comment on this result. Hint: See the solution by γ = 0 using α  OLS and γ Farebrother (1991).

26. Degeneration of Feasible GLS to 2SLS in a Limited Information Simultaneous Equations Model. This is based on Gao and Lahiri (2000). Consider a simple limited information simultaneous equations model, y1

= γy2 + u,

(1)

y2

= Xβ + v,

(2)

where y1 , y2 are N × 1 vectors of observations on two endogenous variables. X is N × K matrix of predetermined variables of the system, and K ≥ 1 such that (1) is identified. Each row of (u, v) is assumed to be i.i.d. (0, Σ), and Σ is p.d. In this case, γ 2SLS = (y2′ PX y2 )−1 y2′ PX y1 , where PX = X(X ′ X)−1 X ′ . The residuals u  = y1 − γ 2SLS y2 and v = M y2 , where M = IN − PX are used to generate a consistent estimate for Σ

′  u ′ v u u = 1 Σ .  v′ v N v′ u

 degenerates to γ Show that a feasible GLS estimate of γ using Σ 2SLS .


References

287

27. Equality of Two IV Estimators. This is based on Qian (1999). Consider the following linear regression model: yi = x′i β + ǫ = x′1i β 1 + x′2i β 2 + ǫi ,

i = 1, 2, . . . , N,

(1)

where the dimensions of x′i , x′1i and x′2i are 1×K, 1×K1 and 1×K2 , respectively, with K = K1 +K2 . xi may be correlated with ǫi , but we have instruments zi such that E(ǫi |zi′ ) = 0 and E(ǫ2i |zi′ ) = σ 2 . ′ ′ ), where the dimensions of zi , z1i , and z2i , z2i Partition the instruments into two subsets: zi′ = (z1i are L, L1 and L2 , with L = L1 + L2 . Assume that E(z1i x′i ) has full column rank (so L1 ≥ K); this ensures that β can be estimated consistently using the subset of instruments z1i only or using the ′ ′ , z2i ). We also assume that (yi , x′i , zi′ ) is covariance stationary. entire set zi′ = (z1i Define PA = A(A′ A)−1 A′ and MA = I − PA for any matrix A with full column rank. Let X =  = (X  −1 X  ′ y,  = P[Z ] X and β  ′ X) (x1 , . . . , xN )′ , and similarly for X1 , X2 , y, Z1 , Z2 and Z. Define X 1  so that β is the instrumental variables (IV) estimator of (1) using Z1 as instruments. Similarly,  = (X  is the IV estimator of (1) using Z as instruments.  = PZ X and β  ′ X)  −1 X  ′ y, so that β define X  =β  if Z ′ M[Z ] [X1 − X2 (X ′ P1 X2 )−1 X ′ P1 X1 ] = 0. Show that β 1 1 2 2 2 1

28. For the crime in North Carolina example given in section 11.6, replicate the results in Tables 11.2-11.6 using the data for 1987. Do the same using the data for 1981. Are there any notable differences in the results as we compare 1981 to 1987?

References This chapter is influenced by Johnston (1984), Kelejian and Oates (1989), Maddala (1992), Davidson and MacKinnon (1993), Mariano (2001) and Stock and Watson (2003). Additional references include the econometric texts cited in Chapter 3 and the following: Anderson, T.W. and H. Rubin (1950), “The Asymptotic Properties of Estimates of the Parameters of a Single Equation in a Complete System of Stochastic Equations,” Annals of Mathematical Statistics, 21: 570-582. Baltagi, B.H. (1989), “A Hausman Specification Test in a Simultaneous Equations Model,” Econometric Theory, Solution 88.3.5, 5: 453-467. Basmann, R.L. (1957), “A Generalized Classical Method of Linear Estimation of Coefficients in a Structural Equation,” Econometrica, 25: 77-83. Basmann, R.L. (1960), “On Finite Sample Distributions of Generalized Classical Linear Identifiability Tests Statistics,” Journal of the American Statistical Association, 55: 650-659. Bekker, P.A. (1994), “Alternative Approximations to the Distribution of Instrumental Variable Estimators,” Econometrica 62: 657-681. Bekker, P.A. and T.J. Wansbeek (2001), “Identification in Parametric Models,” Chapter 7 in Baltagi, B.H. (ed.) A Companion to Theoretical Econometrics (Blackwell: Massachusetts). Bound, J., D.A. Jaeger and R.M. Baker (1995), “Problems with Instrumental Variables Estimation When the Correlation Between the Instruments and the Exogenous Explanatory Variable is Weak,” Journal of American Statistical Association 90: 443-450. Cornwell, C. and W.N. Trumbull (1994), “Estimating the Economic Model of Crime Using Panel Data,” Review of Economics and Statistics, 76: 360-366.


288

Chapter 11: Simultaneous Equations Model

Cragg, J.G. and S.G. Donald (1996), “Inferring the Rank of Matrix,” Journal of Econometrics, 76: 223-250. Deaton, A. (1997), The Analysis of Household Surveys: A Microeconometric Approach to Development Policy (Johns Hopkins University Press: Baltimore). Durbin, J. (1954), “Errors in Variables,” Review of the International Statistical Institute, 22: 23-32. Farebrother, R.W. (1985), “The Exact Bias of Wald’s Estimator,” Econometric Theory, Problem 85.3.1, 1: 419. Farebrother, R.W. (1991), “Comparison of t-Ratios,” Econometric Theory, Solution 90.1.4, 7: 145-146. Fisher, F.M. (1966), The Identification Problem in Econometrics (McGraw-Hill: New York). Gao, C. and K. Lahiri (2000), “Degeneration of Feasible GLS to 2SLS in a Limited Information Simultaneous Equations Model,” Econometric Theory, Problem 00.2.1, 16: 287. Gronau, R. (1973), “The Effects of Children on the Housewife’s Value of Time,” Journal of Political Economy, 81: S168-199. Haavelmo, T. (1944), “The Probability Approach in Econometrics,” Supplement to Econometrica, 12. Hall, A. (1993), “Some Aspects of Generalized Method of Moments Estimation,” Chapter 15 in Handbook of Statistics, Vol. 11 (North Holland: Amsterdam). Hansen, L. (1982), “Large Sample Properties of Generalized Method of Moments Estimators,” Econometrica, 50: 646-660. Hausman, J.A. (1978), “Specification Tests in Econometrics,” Econometrica, 46: 1251-1272. Hausman, J.A. (1983), “Specification and Estimation of Simultaneous Equation Models,” Chapter 7 in Griliches, Z. and Intriligator, M.D. (eds.) Handbook of Econometrics, Vol. I (North Holland: Amsterdam). Holly, A. (1987), “Identification and Estimation of a Simple Two-Equation Model,” Econometric Theory, Problem 87.3.3, 3: 463-466. Holly, A. (1988), “A Hausman Specification Test in a Simultaneous Equations Model,” Econometric Theory, Problem 88.3.5, 4: 537-538. Holly, A. (1990), ”Comparison of t-ratios,” Econometric Theory, Problem 90.1.4, 6: 114. Kapteyn, A. and D.G. Fiebig (1981), “When are Two-Stage and Three-Stage Least Squares Estimators Identical?,” Economics Letters, 8: 53-57. Koopmans, T.C. and J. Marschak (1950), Statistical Inference in Dynamic Economic Models (John Wiley and Sons: New York). Koopmans, T.C. and W.C. Hood (1953), Studies in Econometric Method (John Wiley and Sons: New York). Laffer, A.B., (1970), “Trade Credit and the Money Market,” Journal of Political Economy, 78: 239-267. Lott, W.F. and S.C. Ray (1992), Applied Econometrics: Problems with Data Sets (The Dryden Press: New York). Manski, C.F. (1995), Identification Problems in the Social Sciences (Harvard University Press: Cambridge) Mariano, R.S. (2001), “Simultaneous Equation Model Estimators: Statistical Properties and Practical Implications,” Chapter 6 in Baltagi, B.H. (ed.) A Companion to Theoretical Econometrics (Blackwell: Massachusetts).


Appendix

289

Nelson, C.R. and R. Startz (1990), “The Distribution of the Instrumental Variables Estimator and its t-Ratio when the Instrument is a Poor One,” Journal of Business, 63: S125-140. Sapra, S.K. (1997), “Equivariance of an Instrumental Variable (IV) Estimator in the Linear Regression Model,” Econometric Theory, Problem 97.2.5, 13: 464. Singh, N. and A N. Bhat (1988), “Identification and Estimation of a Simple Two-Equation Model,” Econometric Theory, Solution 87.3.3, 4: 542-545. Staiger, D. and J. Stock (1997), “Instrumental Variables Regression With Weak Instruments,” Econometrica, 65: 557-586. Stock, J.H. and M.W. Watson (2003), Introduction to Econometrics (Addison Wesley: Boston). Theil, H. (1953), “Repeated Least Squares Applied to Complete Equation Systems,” The Hague, Central Planning Bureau (Mimeo). Theil, H. (1971), Principles of Econometrics (Wiley: New York). Wooldridge, J.M. (1990), “A Note on the Lagrange Multiplier and F Statistics for Two Stage Least Squares Regression,” Economics Letters, 34: 151-155. Wald, A. (1940), “Fitting of Straight Lines if Both Variables are Subject to Error,” Annals of Mathematical Statistics, 11: 284-300. Wu, D.M. (1973), “Alternative Tests of Independence Between Stochastic Regressors and Disturbances,” Econometrica, 41: 733-740. Zellner, A. and Theil, H. (1962), “Three-Stage Least Squares: Simultaneous Estimation of Simultaneous Equations,” Econometrica, 30: 54-78.

Appendix: The Identification Problem Revisited: The Rank Condition of Identification In section 11.1.2, we developed a necessary but not sufficient condition for identification. In this section we emphasize that model identification is crucial because only then can we get meaningful estimates of the parameters. For an under-identified model, different sets of parameter values agree well with the statistical evidence. As Bekker and Wansbeek (2001, p. 144) put it, preference for one set of parameter values over other ones becomes arbitrary. Therefore, “Scientific conclusions drawn on the basis of such arbitrariness are in the best case void and in the worst case dangerous.” Manski (1995, p. 6) also warns that “negative identification findings imply that statistical inference is fruitless. It makes no sense to try to use a sample of finite size to infer something that could not be learned even if a sample of infinite size were available.” Consider the simultaneous equation model Byt + Γxt = ut

t = 1, 2, . . . , T.

(A.1)

which displays the whole set of equations at time t. B is G × G, Γ is G × K and ut is G × 1. B is square and nonsingular indicating that the system is complete, i.e., there are as many equations as there are endogenous variables. Premultiplying (A.1) by B −1 and solving for yt in terms of the exogenous variables and the vector of errors, we get yt = Πxt + vt

t = 1, 2, . . . , T.

(A.2)


290

Chapter 11: Simultaneous Equations Model

where Π = −B −1 Γ, is G × K, and vt = B −1 ut . Note that if we premultiply the structural model in (A.1) by an arbitrary nonsingular G × G matrix F , then the new structural model has the same reduced form given in (A.2). In this case, each new structural equation is a linear combination of the original structural equations, but the reduced form equations are the same. One idea for identification, explored by Fisher (1966), is to note that (A.1) is completely defined by B, Γ and the probability density function of the disturbances p(ut ). The specification of the structural model which comes from economic theory, imposes a lot of zero restrictions on the B and Γ coefficients. In addition, there may be cross-equations or within equation restrictions. For example, constant returns to scale of the production function, or homogeneity of a demand equation, or symmetry conditions. In addition, the probability density function of the disturbances may itself contain some zero covariance restrictions. The structural model given in (A.1) is identified if these restrictions are enough to distinguish it from any other structural model. This is operationalized by proving that the only nonsingular matrix F which results in a new structural model that satisfies the same restrictions on the original model is the identity matrix. If after imposing the restrictions, only certain rows of F resemble the corresponding rows of an identity matrix, up to a scalar of normalization, then the corresponding equations of the system are identified. The remaining equations are not identified. This is the same concept of taking a linear combination of the demand and supply equations and seeing if the linear combination is different from demand or supply. If it is, then both equations are identified. If it looks like demand but not supply, then supply is identified and demand is not identified. Let us look at an example. Example (A.1): Consider a demand and supply equations with Qt = a − bPt + cYt + u1t

Qt = d + ePt + f Wt + u2t

(A.3) (A.4)

where Y is income and W is weather. Writing (A.3) and (A.4) in matrix form (A.1), we get

  1 b −a −c 0 Qt B = Γ= yt = (A.5) Pt 1 −e −d 0 −f x′t = [1, Yt , Wt ]

u′t = (u1t , u2t )

There are two zero restrictions on Γ. The first is that income does not appear in the supply equation and the second is that weather does not appear in the demand equation. Therefore, the order condition of identification is satisfied for both equations. In fact, for each equation, there is one excluded exogenous variable and only one right hand side included endogenous variable. Therefore, both equations are just-identified. Let F = [fij ] for i, j = 1, 2, be a nonsingular matrix. Premultiply this system by F . The new matrix B is now F B and the new matrix Γ is now F Γ. In order for the transformed system to satisfy the same restrictions as the original model, F B must satisfy the following normalization restrictions: f11 + f12 = 1

f21 + f22 = 1

(A.6)

Also, F Γ should satisfy the following zero restrictions −f21 c + f22 0 = 0

f11 0 − f12 f = 0

(A.7)

Since c = 0, and f = 0, then (A.7) implies that f21 = f12 = 0. Using (A.6), we get f11 = f22 = 1. Hence, the only nonsingular F that satisfies the same restrictions on the original model is the identity matrix, provided c = 0 and f = 0. Therefore, both equations are identified.


Appendix

291

Example (A.2): If income does not appear in the demand equation (A.3), i.e., c = 0, the model looks like Qt = a − bPt + u1t

Qt = d + ePt + f Wt + u2t

(A.8) (A.9)

In this case, only the second restriction given in (A.7) holds. Therefore, f = 0 implies f12 = 0, however f21 is not necessarily zero without additional restrictions. Using (A.6), we get f11 = 1 and f21 + f22 = 1. This means that only the first row of F looks like the first row of an identity matrix, and only the demand equation is identified. In fact, the order condition for identification is not met for the supply equation since there are no excluded exogenous variables from that equation but there is one right hand side included endogenous variable. See problems 6 and 7 for more examples of this method of identification. Example (A.3): Suppose that u ∼ (0, Ω) where Ω = Σ ⊗ IT , and Σ = [σ ij ] for i, j = 1, 2. This example shows how a variance-covariance restriction can help identify an equation. Let us take the model defined in (A.8), (A.9) and add the restriction that σ 12 = σ 21 = 0. In this case, the transformed model disturbances will be F u ∼ (0, Ω∗ ), where Ω∗ = Σ∗ ⊗ IT , and Σ∗ = F ΣF ′ . In fact, since Σ is diagonal, F ΣF ′ should also be diagonal. This imposes the following restriction on the elements of F : f11 σ 11 f21 + f12 σ 22 f22 = 0

(A.10)

But, f11 = 1 and f12 = 0 from the zero restrictions imposed on the demand equation, see example 4. Hence, (A.10) reduces to σ 11 f21 = 0. Since σ 11 = 0, this implies that f21 = 0, and the normalization restriction given in (A.6), implies that f22 = 1. Therefore, the second equation is also identified. Example (A.4): In this example, we demonstrate how cross-equation restrictions can help identify equations. Consider the following simultaneous model y1 = a + by2 + cx1 + u1

(A.11)

y2 = d + ey1 + f x1 + gx2 + u2

(A.12)

and add the restriction c = f . It can be easily shown that the first equation is identified with f11 = 1, and f12 = 0. The second equation has no zero restrictions, but the cross-equation restriction c = f implies: −cf11 − f f12 = −cf21 − f f22 Using c = f , we get f11 + f12 = f21 + f22

(A.13)

But, the first equation is identified with f11 = 1 and f12 = 0. Hence, (A.13) reduces to f21 +f22 = 1, which together with the normalization condition −f21 b + f22 = 1, gives f21 (b + 1) = 0. If b = −1, then f21 = 0 and f22 = 1. The second equation is also identified provided b = −1. Alternatively, one can look at the problem of identification by asking whether the structural parameters in B and Γ can be obtained from the reduced form parameters. It will be clear from the following discussion that this task is impossible if there are no restrictions on this


292

Chapter 11: Simultaneous Equations Model

simultaneous model. In this case, the model is hopelessly unidentified. However, in the usual case where there are a lot of zeros in B and Γ, we may be able to retrieve the remaining non-zero coefficients from Π. More rigorously, Π = −B −1 Γ, which implies that BΠ + Γ = 0

(A.14)

or AW = 0

where A = [B, Γ]

and W ′ = [Π′ , IK ]

(A.15)

For the first equation, this implies that α′1 W = 0 where α′1 is the first row of A.

(A.16)

W is known (or can be estimated) and is of rank K. If the first equation has no restrictions on its structural parameters, then α′1 contains (G + K) unknown coefficients. These coefficients satisfy K homogeneous equations given in (A.16). Without further restrictions, we cannot solve for (G + K) coefficients (α′1 ) with only K equations. Let φ denote the matrix of R zero restrictions on the first equation, i.e., α′1 φ = 0. This together with (A.16) implies that α′1 [W, φ] = 0

(A.17)

and we can solve uniquely for α′1 (up to a scalar of normalization) provided the rank [W, φ] = G + K − 1

(A.18)

Economists specify each structural equation with the left hand side endogenous variable having the coefficient one. This normalization identifies one coefficient of α′1 , therefore, we require only (G + K − 1) more restrictions to uniquely identify the remaining coefficients of α1 . [W, φ] is a (G + K) × (K + R) matrix. Its rank is less than any of its two dimensions, i.e., (K + R) ≥ (G + K − 1), which results in R ≥ G − 1, or the order condition of identification. Note that this is a necessary but not sufficient condition for (A.18) to hold. It states that the number of restrictions on the first equation must be greater than the number of endogenous variables minus one. If all R restrictions are zero restrictions, then it means that the number of excluded exogenous plus the number of excluded endogenous variables should be greater than (G − 1). But the G endogenous variables are made up of the left hand side endogenous variable y1 , the g1 right hand side included endogenous variables Y1 , and (G − g1 − 1) excluded endogenous variables. Therefore, R ≥ (G − 1) can be written as k2 + (G − g1 − 1) ≥ (G − 1) which reduces to k2 ≥ g1 , which was discussed earlier in this chapter. The necessary and sufficient condition for identification can now be obtained as follows: Using (A.1) one can write Azt = ut

where zt′ = (yt′ , x′t )

(A.19)

and from the first definition of identification we make the transformation F Azt = F ut , where F is a G × G nonsingular matrix. The first equation satisfies the restrictions α′1 φ = 0, which can be rewritten as ι′ Aφ = 0, where ι′ is the first row of an identity matrix IG . F must satisfy the restriction that (first row of F A)φ = 0. But the first row of F A is the first row of F , say f1′ , times A. This means that f1′ (Aφ) = 0. For the first equation to be identified, this condition


Appendix

293

on the transformed first equation must be equivalent to ι′ Aφ = 0, up to a scalar constant. This holds if and only if f1′ is a scalar multiple of ι′ , and the latter condition holds if and only if the rank (Aφ) = G − 1. The latter is known as the rank condition for identification. Example (A.5): Consider the simple Keynesian model given in example 1. The second equation is an identity and the first equation satisfies the order condition of identification, since It is the excluded exogenous variable from that equation, and there is only one right hand side included endogenous variable Yt . In fact, the first equation is just-identified. Note that

β 11 β 12 γ 11 γ 12 1 −β −α 0 A = [B, Γ] = = (A.20) β 21 β 22 γ 21 γ 22 −1 1 0 −1 and φ for the first equation consists of only one restriction, namely that It is not in that equation, or γ 12 = 0. This makes φ′ = (0, 0, 0, 1), since α′1 φ = 0 gives γ 12 = 0. From (A.20), Aφ = (γ 12 , γ 22 )′ = (0, −1)′ and the rank (Aφ) = 1 = G − 1. Hence, the rank condition holds for the first equation and it is identified. Problem 8 reconsiders example (A.1), where both equations are just-identified by the order condition of identification and asks the reader to show that both satisfy the rank condition of identification as long as c = 0 and f = 0. The reduced form of the simple Keynesian model is given in equations (11.4) and (11.5). In fact,

α β π 11 π 12 = /(1 − β (A.21) Π= π 21 π 22 α 1 Note that π 11 /π 22 = α and

π 21 /π 22 = α

π 12 /π 22 = β

(π 22 − 1)/π 22 = β

and

(A.22)

Therefore, the structural parameters of the consumption equation can be retrieved from the  OLS ? reduced form coefficients. However, what happens if we replace Π by its OLS estimate Π Would the solution in (A.22) lead to two estimates of (α, β) or would this solution lead to a unique estimate? In this case, the consumption equation is just-identified and the solution in (A.22) is unique. To show this, recall that π 12 = mci /mii ;

and π 22 = myi /mii

 using (A.22), one gets Solving for β, and

(A.23)

=π β 12 / π 22 = mci /myi

(A.24)

 = ( β π 22 − 1)/ π 22 = (mci − myi )/myi

(A.25)

myi = mci + mii

(A.26)

(A.24) and (A.25) lead to a unique solution because equation (11.2) gives

In general, we would not be able to solve for the structural parameters of an unidentified equation in terms of the reduced form parameters. However, when this equation is identified,


294

Chapter 11: Simultaneous Equations Model

replacing the reduced form parameters by their OLS estimates would lead to a unique estimate of the structural parameters, only if this equation is just-identified, and to more than one estimate depending upon the degree of over-identification. Problem 8 gives another example of the just-identified case, while problem 9 considers a model with one unidentified and another over-identified equation. Example (A.6): Equations (11.13) and (11.14) give an unidentified demand and supply model with

−α 1 −β (A.27) and Γ = B= −γ 1 −δ The reduced form equations given by (11.16) and (11.17) yield

αδ − γβ π 11 −1 /(δ − β) = Π = −B Γ = π 21 α−γ

(A.28)

Note that one cannot solve for (α, β) nor (γ, δ) in terms of (π 11 , π 21 ) without further restrictions.


CHAPTER 12

Pooling Time-Series of Cross-Section Data 12.1

Introduction

In this chapter, we will consider pooling time-series of cross-sections. This may be a panel of households or firms or simply countries or states followed over time. Two well known examples of panel data in the U.S. are the Panel Study of Income Dynamics (PSID) and the National Longitudinal Survey (NLS). The PSID began in 1968 with 4802 families, including an over-sampling of poor households. Annual interviews were conducted and socioeconomic characteristics of each of the families and of roughly 31000 individuals who have been in these or derivative families were recorded. The list of variables collected is over 5000. The NLS, followed five distinct segments of the labor force. The original samples include 5020 older men, 5225 young men, 5083 mature women, 5159 young women and 12686 youths. There was an over-sampling of blacks, hispanics, poor whites and military in the youths survey. The list of variables collected runs into the thousands. An inventory of national studies using panel data is given at http://www.isr.umich.edu/src/psid/panelstudies.html. Pooling this data gives a richer source of variation which allows for more efficient estimation of the parameters. With additional, more informative data, one can get more reliable estimates and test more sophisticated behavioral models with less restrictive assumptions. Another advantage of panel data sets are their ability to control for individual heterogeneity. Not controlling for these unobserved individual specific effects leads to bias in the resulting estimates. Panel data sets are also better able to identify and estimate effects that are simply not detectable in pure cross-sections or pure timeseries data. In particular, panel data sets are better able to study complex issues of dynamic behavior. For example, with a cross-section data set one can estimate the rate of unemployment at a particular point in time. Repeated cross-sections can show how this proportion changes over time. Only panel data sets can estimate what proportion of those who are unemployed in one period remain unemployed in another period. Some of the benefits and limitations of using panel data sets are listed in Hsiao (2003) and Baltagi (2005). Section 12.2 studies the error components model focusing on fixed effects, random effects and maximum likelihood estimation. Section 12.3 considers the question of prediction in a random effects model, while Section 12.4 illustrates the estimation methods using an empirical example. Section 12.5 considers testing the poolability assumption, the existence of random individual effects and the consistency of the random effects estimator using a Hausman test. Section 12.6 studies the dynamic panel data model and illustrates the methods used with an empirical example. Section 12.7 concludes with a short presentation of program evaluation and the difference-in-differences estimator.

12.2

The Error Components Model

The regression model is still the same, but it now has double subscripts yit = α + Xit′ β + uit

(12.1)


296

Chapter 12: Pooling Time-Series of Cross-Section Data

where i denotes cross-sections and t denotes time-periods with i = 1, 2, . . . , N , and t = 1, 2, . . . , T. α is a scalar, β is K × 1 and Xit is the it-th observation on K explanatory variables. The observations are usually stacked with i being the slower index, i.e., the T observations on the first household followed by the T observations on the second household, and so on, until we get to the N -th household. Under the error components specification, the disturbances take the form uit = μi + ν it

(12.2)

where the μi ’s are cross-section specific components and ν it are remainder effects. For example, μi may denote individual ability in an earnings equation, or managerial skill in a production function or simply a country specific effect. These effects are time-invariant. In vector form, (12.1) can be written as y = αιN T + Xβ + u = Zδ + u

(12.3)

where y is N T × 1, X is N T × K, Z = [ιN T , X], δ ′ = (α′ , β ′ ), and ιN T is a vector of ones of dimension N T . Also, (12.2) can be written as u = Zμ μ + ν

(12.4)

where u′ = (u11 , . . . , u1T , u21 , . . . , u2T , . . . , uN 1 , . . . , uN T ) and Zμ = IN ⊗ ιT . IN is an identity matrix of dimension N, ιT is a vector of ones of dimension T , and ⊗ denotes Kronecker product defined in the Appendix to Chapter 7. Zμ is a selector matrix of ones and zeros, or simply the matrix of individual dummies that one may include in the regression to estimate the μi ’s if they are assumed to be fixed parameters. μ′ = (μ1 , . . . , μN ) and ν ′ = (ν 11 , . . . , ν 1T , . . . , ν N 1 , . . . , ν N T ). Note that Zμ Zμ′ = IN ⊗JT where JT is a matrix of ones of dimension T , and P = Zμ (Zμ′ Zμ )−1 Zμ′ , the projection matrix on Zμ , reduces to P = IN ⊗J¯T where J¯T = JT /T . P is a matrix which averages the observation across time for each individual, and Q = IN T − P is a matrix which T obtains the deviations from individual means. For example, P u has a typical element u ¯i. = t=1 uit /T repeated T times for each individual and Qu has a typical element (uit − u ¯i. ). P and Q are (i) symmetric idempotent matrices, i.e., P ′ = P and P 2 = P . This means that the rank (P ) = tr(P ) = N and rank (Q) = tr(Q) = N (T − 1). This uses the result that rank of an idempotent matrix is equal to its trace, see Graybill (1961, Theorem 1.63) and the Appendix to Chapter 7. Also, (ii) P and Q are orthogonal, i.e., P Q = 0 and (iii) they sum to the identity matrix P + Q = IN T . In fact, any two of these properties imply the third, see Graybill (1961, Theorem 1.68). 12.2.1

The Fixed Effects Model

If the μi ’s are thought of as fixed parameters to be estimated, then equation (12.1) becomes  yit = α + Xit′ β + N (12.5) i=1 μi Di + ν it

where Di is a dummy variable for the i-th household. Not all the dummies are included so as not to fall in the dummy variable trap. One Nis usually dropped or equivalently, we can say that there is a restriction on the μ’s given by i=1 μi = 0. The ν it ’s are the usual classical IID random variables with 0 mean and variance σ 2ν . OLS on equation (12.5) is BLUE, but we have two


12.2

The Error Components Model

297

problems, the first is the loss of degrees of freedom since in this case, we are estimating N + K parameters. Also, with a lot of dummies we could be running into multicollinearity problems and a large X ′ X matrix to invert. For example, if N = 50 states, T = 10 years and we have two explanatory variables, then with 500 observations we are estimating 52 parameters. Alternatively, we can think of this in an analysis of variance context and rearrange our observations, say, on y in an (N × T ) matrix where rows denote firms and columns denote time periods.

i

1 2 : N

1 y11 y21 : yN 1

2 y12 y22 : yN 2

t .. .. .. .. ..

T y1T y2T : yN T

y1. y2. : yN.

 where yi. = Tt=1 yit and y¯i. = yi. /T . For the simple regression with one regressor, the model given in (12.1) becomes yit = α + βxit + μi + ν it

(12.6)

averaging over time gives y¯i. = α + β x ¯i. + μi + ν¯i.

(12.7)

and averaging over all observations gives y¯.. = α + β x ¯.. + ν¯.. (12.8)  T where y¯.. = N t=1 yit /N T . Equation (12.8) follows because the μi ’s sum to zero. Defining i=1 yit = (yit − y¯i. ) and x it and  ν it similarly, we get ¯i. ) + (ν it − ν¯i. ) yit − y¯i. = β(xit − x

or xit +  ν it yit = β

(12.9)

Running OLS on equation (12.9) leads to the same estimator of β as that obtained from equation  in our notation. (12.5). This is called the least squares dummy  variable estimator (LSDV) or β N T 2 it is the within sum of squares in an It is also known as the Within estimator since i=1 t=1 x analysis of variance framework. One can then retrieve an estimate of α from equation (12.8) as x α  = y¯.. − β ¯.. . Similarly, if we are interested in the μi ’s, those can also be retrieved from (12.7) and (12.8) as follows:  xi. − x μ i = (¯ yi. − y¯.. ) − β(¯ ¯.. )

(12.10)

y = αιN T + Xβ + Zμ μ + ν = Zδ + Zμ μ + ν

(12.11)

In matrix form, one can substitute the disturbances given by (12.4) into (12.3) to get

and then perform OLS on (12.11) to get estimates of α, β and μ. Note that Z is N T × (K + 1) and Zμ , the matrix of individual dummies is N T × N . If N is large, (12.11) will include too


298

Chapter 12: Pooling Time-Series of Cross-Section Data

many individual dummies, and the matrix to be inverted by OLS is large and of dimension (N + K). In fact, since α and β are the parameters of interest, one can obtain the least squares dummy variables (LSDV) estimator from (12.11), by residualing out the Zμ variables, i.e., by premultiplying the model by Q, the orthogonal projection of Zμ , and performing OLS Qy = QXβ + Qν

(12.12)

This uses the fact that QZμ = QιN T = 0, since P Zμ = Zμ . In other words, the Q matrix wipes out the individual effects. Recall, the FWL Theorem in Chapter 7. This is a regression  = QX with typical element (Xit,k − X ¯ i.,k ) for of y = Qy with typical element (yit − y¯i. ) on X the k-th regressor, k = 1, 2, . . . , K. This involves the inversion of a (K × K) matrix rather than (N + K) × (N + K) as in (12.11). The resulting OLS estimator is  = (X ′ QX)−1 X ′ Qy β

(12.13)

 = σ 2 (X ′ QX)−1 = σ 2 (X  ′ X)  −1 . with var(β) ν ν Note that this fixed effects (FE) estimator cannot estimate the effect of any time-invariant variable like sex, race, religion, schooling, or union participation. These time-invariant variables are wiped out by the Q transformation, the deviations from means transformation. Alternatively, one can see that these time-invariant variables are spanned by the individual dummies in (12.5) and therefore any regression package attempting (12.5) will fail, signaling perfect multicollinearity. If (12.5) is the true model, LSDV is BLUE as long as ν it is the standard classical disturbance with mean 0 and variance covariance matrix σ 2ν IN T . Note that as T → ∞, the FE estimator is consistent. However, if T is fixed and N → ∞ as typical in short labor panels, then only the FE estimator of β is consistent, the FE estimators of the individual effects (α + μi ) are not consistent since the number of these parameters increase as N increases. Testing for Fixed Effects: One could test the joint significance of these dummies, i.e., H0 ; μ1 = μ2 = .. = μN −1 = 0, by performing an F -test. This is a simple Chow test given in (4.17) with the restricted residual sums of squares (RRSS) being that of OLS on the pooled model and the unrestricted residual sums of squares (URSS) being that of the LSDV regression. If N is large, one can perform the within transformation and use that residual sum of squares as the URSS. In this case F0 =

(RRSS − URSS )/(N − 1) H0 ∼ FN−1,N(T −1)−K URSS /(N T − N − K)

(12.14)

Computational Warning: One computational caution for those using the Within regression given by (12.12). The s2 of this regression as obtained from a typical regression package divides the residual sums of squares by N T − K since the intercept and the dummies are not included. The proper s2 , say s∗2 from the LSDV regression in (12.5) would divide the same residual sums of squares by N (T − 1) − K. Therefore, one has to adjust the variances obtained from the within regression (12.12) by multiplying the variance-covariance matrix by (s∗2 /s2 ) or simply by multiplying by [N T − K]/[N (T − 1) − K]. 12.2.2

The Random Effects Model

There are too many parameters in the fixed effects model and the loss of degrees of freedom can be avoided if the μi ’s can be assumed random. In this case μi ∼ IID(0, σ 2μ ), ν it ∼ IID(0, σ 2ν )


12.2

The Error Components Model

299

and the μi ’s are independent of the ν it ’s. In addition, the Xit ’s are independent of the μi ’s and ν it ’s for all i and t. The random effects model is an appropriate specification if we are drawing N individuals randomly from a large population. This specification implies a homoskedastic variance var(uit ) = σ 2μ + σ 2ν for all i and t, and an equi-correlated block-diagonal covariance matrix which exhibits serial correlation over time only between the disturbances of the same individual. In fact, cov(uit , ujs ) = σ 2μ + σ 2ν =

σ 2μ

for i = j, t = s for

(12.15)

i = j, t = s

and zero otherwise. This also means that the correlation coefficient between uit and ujs is ρ = correl(uit , ujs ) = 1 =

σ 2μ /(σ 2μ

+

σ 2ν )

for i = j, t = s for

(12.16)

i = j, t = s

and zero otherwise. From (12.4), one can compute the variance-covariance matrix Ω = E(uu′ ) = Zμ E(μμ′ )Zμ′ + E(νν ′ ) = σ 2μ (IN ⊗ JT ) + σ 2ν (IN ⊗ IT )

(12.17)

In order to obtain the GLS estimator of the regression coefficients, we need Ω−1 . This is a huge matrix for typical panels and is of dimension (N T × N T ). No brute force inversion should be attempted even if the researcher’s application has a small N and T . For example, if we observe N = 20 firms over T = 5 time periods, Ω will be 100 by 100. We will follow a simple trick devised by Wansbeek and Kapteyn (1982) that allows the deviation of Ω−1 and Ω−1/2 . Essentially, one replaces JT by T J¯T , and IT by (ET + J¯T ) where ET is by definition (IT − J¯T ). In this case: Ω = T σ 2μ (IN ⊗ J¯T ) + σ 2ν (IN ⊗ ET ) + σ 2ν (IN ⊗ J¯T ) collecting terms with the same matrices, we get Ω = (T σ 2μ + σ 2ν )(IN ⊗ J¯T ) + σ 2ν (IN ⊗ ET ) = σ 21 P + σ 2ν Q

(12.18)

where σ 21 = T σ2μ + σ 2ν . (12.18) is the spectral decomposition representation of Ω, with σ 21 being the first unique characteristic root of Ω of multiplicity N and σ 2ν is the second unique characteristic root of Ω of multiplicity N (T − 1). It is easy to verify, using the properties of P and Q, that Ω−1 =

1 1 P + 2Q σν σ 21

(12.19)

and Ω−1/2 =

1 1 P+ Q σ1 σν

(12.20)

In fact, Ωr = (σ 21 )r P + (σ 2ν )r Q where r is an arbitrary scalar. Now we can obtain GLS as a weighted least squares. Fuller and Battese (1974) suggested premultiplying the regression equation given in (12.3) by σ ν Ω−1/2 = Q + (σ ν /σ 1 )P and performing OLS on the resulting transformed regression. In this case, y ∗ = σ ν Ω−1/2 y has a typical element yit − θy¯i. where θ = 1 − (σ ν /σ 1 ). This transformed regression inverts a matrix of dimension (K + 1) and can be easily implemented using any regression package.


300

Chapter 12: Pooling Time-Series of Cross-Section Data

The Best Quadratic Unbiased (BQU) estimators of the variance components arise naturally from the spectral decomposition of Ω. In fact, P u ∼ (0, σ 21 P ) and Qu ∼ (0, σ 2ν Q) and

and

σ 21 = σ 2ν =

 u′ P u =T N ¯2i. /N i=1 u tr(P )  T u′ Qu ¯i. )2 /N (T − 1) =T N t=1 (uit − u i=1 tr(Q)

(12.21)

(12.22)

provide the BQU estimators of σ 21 and σ 2ν , respectively, see Balestra (1973). These are analysis of variance type estimators of the variance components and are MVU under normality of the disturbances, see Graybill (1961). The true disturbances are not known and therefore (12.21) and (12.22) are not feasible. Wallace and Hussain (1969) suggest substituting OLS residuals u OLS instead of the true u’s. After all, the OLS estimates are still unbiased and consistent, but no longer efficient. Amemiya (1971) shows that these estimators of the variance components have a different asymptotic distribution from that knowing the true disturbances. He suggests using the LSDV residuals instead of the OLS residuals. In this case u  = y−α  ιN T −  where α  and X ¯ ..′ β ¯ ..′ is a 1×K vector of averages of all regressors. Substituting these Xβ  = y¯.. − X u ’s for u in (12.21) and (12.22) we get the Amemiya-type estimators of the variance components. The resulting estimates of the variance components have the same asymptotic distribution as that knowing the true disturbances. Swamy and Arora (1972) suggest running two regressions to get estimates of the variance components from the corresponding mean square errors of these regressions. The first regression is the Within regression, given in (12.12), which yields the following s2 : 2  σ ν = [y ′ Qy − y ′ QX(X ′ QX)−1 X ′ Qy]/[N (T − 1) − K]

(12.23)

The second regression is the Between regression which runs the regression of averages across time, i.e., ¯′ β + u y¯i. = α + X ¯i. i.

i = 1, . . . , N

(12.24)

This is equivalent to premultiplying the model in (12.11) by P and running OLS. The only caution is that the latter regression has N T observations because it repeats the averages T times for each individual, while the cross-section regression in (12.24) is based on N observations. To remedy this, one can run the cross-section regression √ √ √ √ (12.25) yi. / T = α( T ) + (Xi.′ / T )β + ui. / T √ where one can easily verify that var(ui. / T ) = σ 21 . This regression will yield an s2 given by 2  σ 1 = (y ′ P y − y ′ P Z(Z ′ P Z)−1 Z ′ P y)/(N − K − 1)

(12.26)

Note that stacking the following two transformed regressions we just performed yields       Qy QZ Qu = δ+ (12.27) Py PZ Pu


12.2

The Error Components Model

301

and the transformed error has mean 0 and variance-covariance matrix given by  2  σν Q 0 0 σ 21 P Problem 6 asks the reader to verify that OLS on this system of 2N T observations yields OLS on the pooled model (12.3). Also, GLS on this system yields GLS on (12.3). Alternatively, one could get rid of the constant α by running the following stacked regressions: 

Qy (P − J¯N T )y



=



QX (P − J¯N T )X



β+



Qu (P − J¯N T )u



(12.28)

This follows from the fact the QιN T = 0 and (P − J¯N T )ιN T = 0. The transformed error has zero mean and variance-covariance matrix  2  σν Q 0 (12.29) 0 σ 21 (P − J¯N T ) OLS on this system, yields OLS on (12.3) and GLS on (12.28) yields GLS on (12.3). In fact,  β GLS

= [(X ′ QX/σ 2ν ) + X ′ (P − J¯N T )X/σ 21 ]−1 [(X ′ Qy/σ 2ν ) + (X ′ (P − J¯N T )y/σ 21 )]

= [WXX + φ2 BXX ]−1 [WXy + φ2 BXy ]

(12.30)

2 −1 ′ ′  ¯ with var(β GLS ) = σ ν [WXX + φ BXX ] . Note that WXX = X QX, BXX = X (P − JN T )X −1 2 2 2  and φ = σ ν /σ 1 . Also, the Within estimator of β is β W ithin = WXX WXy and the Between −1    estimator β Between = BXX BXy . This shows that β GLS is a matrix weighted average of β W ithin  and β Between weighing each estimate by the inverse of its corresponding variance. In fact 2

   β GLS = W1 β W ithin + W2 β Between

(12.31)

where W1 = [WXX + φ2 BXX ]−1 WXX and W2 = [WXX + φ2 BXX ]−1 (φ2 BXX ) = I − W1 . This  was demonstrated by Maddala (1971). Note that (i) if σ 2μ = 0, then φ2 = 1 and β GLS reduces 2 2 OLS . (ii) If T → ∞, then φ → 0 and β GLS tends to β  W ithin . (iii) If φ → ∞, then β GLS to β  tends to β . In other words, the Within estimator ignores the between variation, and Between the Between estimator ignores the within variation. The OLS estimator gives equal weight to   the between and within variations. From (12.30), it is clear that var(β W ithin )− var(β GLS ) is a 2 positive semi-definite matrix, since φ is positive. However as T → ∞ for any fixed N , φ2 → 0   and both β GLS and β W ithin have the same asymptotic variance. Another estimator of the was suggested by Nerlove (1971). His sugges variance components 2 /(N − 1) where μ μ ˆ ) i are the dummy coefficients estimates ( μ − tion is to estimate σ 2μ = N i=1 i 2 from the LSDV regression. σ ν is estimated from the within residual sums of squares divided by N T without correction for degrees of freedom. 2ν )/T . In this σ 21 − σ Note that, except for Nerlove’s (1971) method, one has to retrieve σ 2μ as ( 2 case, there is no guarantee that the estimate of σ μ would be non-negative. Searle (1971) has an extensive discussion of the problem of negative estimates of the variance components in the biometrics literature. One solution is to replace these negative estimates by zero. This in fact is the suggestion of the Monte Carlo study by Maddala and Mount (1973). This study finds that negative estimates occurred only when the true σ 2μ was small and close to zero. In these cases


302

Chapter 12: Pooling Time-Series of Cross-Section Data

OLS is still a viable estimator. Therefore, replacing negative σ 2μ by zero is not a bad sin after all, and the problem is dismissed as not being serious. Under the random effects model, GLS based on the true variance components is BLUE, and all the feasible GLS estimators considered are asymptotically efficient as either N or T → ∞. Maddala and Mount (1973) compared OLS, Within, Between, feasible GLS methods, true GLS and MLE using their Monte Carlo study. They found little to choose among the various feasible GLS estimators in small samples and argued in favor of methods that were easier to compute. Taylor (1980) derived exact finite sample results for the one-way error components model. He compared the Within estimator with the Swamy-Arora feasible GLS estimator. He found the following important results: (1) Feasible GLS is more efficient than FE for all but the fewest degrees of freedom. (2) The variance of feasible GLS is never more than 17% above the Cram´erRao lower bound. (3) More efficient estimators of the variance components do not necessarily yield more efficient feasible GLS estimators. These finite sample results are confirmed by the Monte Carlo experiments carried out by Maddala and Mount (1973) and Baltagi (1981). 12.2.3

Maximum Likelihood Estimation

Under normality of the disturbances, one can write the log-likelihood function as L(α, β, φ2 , σ 2ν ) = constant −

N 1 NT logσ 2ν + logφ2 − 2 u′ Σ−1 u 2 2 2σ ν

(12.32)

where Ω = σ 2ν Σ, φ2 = σ 2ν /σ 21 and Σ = Q + φ−2 P from (12.18). This uses the fact that |Ω| = product of its characteristic roots = (σ 2ν )N (T −1) (σ 21 )N = (σ 2ν )N T (φ2 )−N . Note that there is a one-to-one correspondence between φ2 and σ 2μ . In fact, 0 ≤ σ 2μ < ∞ translates into 0 < φ2 ≤ 1. Brute force maximization of (12.32) leads to nonlinear first-order conditions, see Amemiya (1971). Instead, Breusch (1987) concentrates the likelihood with respect to α and σ 2ν . In this  ¯′ β  −1 u  are based on MLE’s of 2ν,M LE = u ′ Σ /N T where u  and Σ case, α  M LE = y¯.. − X .. M LE and σ M LE then α β, φ2 and α. Let d = y − X β  M LE = ι′N T d/N T and u  = d − ιN T α  = d − J¯N T d. This 2 implies that σ ν,M LE can be rewritten as σ 2ν,M LE = d′ [Q + φ2 (P − J¯N T )]d/N T

(12.33)

and the concentrated log-likelihood becomes Lc (β, φ2 ) = constant −

NT N log{d′ [Q + φ2 (P − J¯N T )]d} + logφ2 2 2

Maximizing (12.34), over φ2 given β, yields N T ¯ 2 2 d′ Qd t=1 (dit − di. ) i=1  = φ =  ¯ ¯ 2 (T − 1)d′ (P − J¯N T )d T (T − 1) N i=1 (di. − d.. )

Maximizing (12.34) over β, given φ2 , yields  ′ −1 ′ 2  ¯ X [Q + φ2 (P − J¯N T )]y β M LE = X [Q + φ (P − JN T )]X

(12.34)

(12.35)

(12.36)

One can iterate between β and φ2 until convergence. Breusch (1987) shows that provided T > 1, any i-th iteration β, call it β i , gives 0 < φ2i+1 < ∞ in the (i+1)th iteration. More importantly,


12.3

Prediction

303

Breusch (1987) shows that these φ2i ’s have a remarkable property of forming a monotonic sequence. In fact, starting from the Within estimator of β, for φ2 = 0, the next φ2 is finite and positive and starts a monotonically increasing sequence of φ2 ’s. Similarly, starting from the Between estimator of β, for (φ2 → ∞) the next φ2 is finite and positive and starts a monotonically decreasing sequence of φ2 ’s. Hence, to guard against the possibility of a local maximum,   Breusch (1987) suggests starting with β W ithin and β Between and iterating. If these two sequences  converge to the same maximum, then this is the global maximum. If one starts with β for OLS

φ2 = 1, and the next iteration obtains a larger φ2 , then we have a local maximum at the boundary φ2 = 1. Maddala (1971) finds that there are at most two maxima for the likelihood L(φ2 ) for 0 < φ2 ≤ 1. Hence, we have to guard against one local maximum.

12.3

Prediction

Suppose we want to predict S periods ahead for the i-th individual. For the random effects model, the BLU estimator is GLS. Using the results in Chapter 9 on GLS, Goldberger’s (1962) Best Linear Unbiased Predictor (BLUP) of yi,T +S is ′ ′ −1  GLS yˆi,T +S = Zi,T +S δ GLS + w Ω u

for S ≥ 1

where u GLS = y − Z  δGLS and w = E(ui,T +S u). Note that ui,T +S = μi + ν i,T +S

(12.37)

(12.38)

and w = σ 2μ (ℓi ⊗ ιT ) where ℓi is the i-th column of IN , i.e., ℓi is a vector that has 1 in the i-th position and zero elsewhere. In this case

σ 2μ 1 1 ′ ′ −1 2 ′ (12.39) w Ω = σ μ (ℓi ⊗ ιT ) 2 P + 2 Q = 2 (ℓ′i ⊗ ι′T ) σν σ1 σ1 since (ℓ′i ⊗ ι′T ) P = (ℓ′i ⊗ ι′T ) and (ℓ′i ⊗ ι′T )Q = 0. The typical element of w′ Ω−1 u GLS is    (T σ 2μ /σ 21 )u ¯i.,GLS where u ¯i.,GLS = Tt=1 u it,GLS /T . Therefore, in (12.37), the BLUP for yi,T +S corrects the GLS prediction by a fraction of the mean of the GLS residuals corresponding to that i-th individual. This predictor was considered by Wansbeek and Kapteyn (1978) and Taub (1979).

12.4

Empirical Example

Baltagi and Griffin (1983) considered the following gasoline demand equation: log

Gas Y PM G Car = α + β 1 log + β 2 log + β 3 log +u Car N PGDP N

(12.40)

where Gas/Car is motor gasoline consumption per auto, Y/N is real income per capita, PM G / PGDP is real motor gasoline price and Car/N denotes the stock of cars per capita. This panel consists of annual observations across eighteen OECD countries, covering the period 1960-1978. The data for this example are provided on the Springer web site as GASOLINE.DAT. Table 12.1


304

Chapter 12: Pooling Time-Series of Cross-Section Data

gives the Stata output for the Within estimator using xtreg, f e. This is the regression described in (12.5) and computed as in (12.9). The Within estimator gives a low price elasticity for gasoline demand of -.322. The F -statistic for the significance of the country effects described in (12.14) yields an observed value of 83.96. This is distributed under the null as an F (17, 321) and is statistically significant. This F -statistic is printed by Stata below the fixed effects output. In EViews, one invokes the test for redundant effects after running the fixed effects regression. Table 12.1 Fixed Effects Estimator – Gasoline Demand Data Coef.

Std. Err.

T

P > |t|

log(Y /N ) log(PMG /PGDP ) log(Car/N ) Constant

0.6622498 –0.3217025 –0.6404829 2.40267

0.073386 0.0440992 0.0296788 0.2253094

9.02 –7.29 –21.58 10.66

sigma u sigma e Rho

0.34841289 0.09233034 0.93438173

(fraction of variance due to u i)

0.000 0.000 0.000 0.000

[95% Conf. Interval] 0.5178715 –0.4084626 –0.6988725 1.959401

0.8066282 –0.2349425 –0.5820933 2.84594

Table 12.2 gives the Stata output for the Between estimator using xtreg, be.This is based on the regression given in (12.24). The Between estimator yields a high price elasticity of gasoline demand of -.964. These results were also verified using TSP. Table 12.2 Between Estimator – Gasoline Demand Data

log(Y /N ) log(PMG /PGDP ) log(Car/N ) Constant

Coef.

Std. Err.

T

0.9675763 –0.9635503 –0.795299 2.54163

0.1556662 0.1329214 0.0824742 0.5267845

6.22 –7.25 –9.64 4.82

P > |t| 0.000 0.000 0.000 0.000

[95% Conf. Interval] 0.6337055 –1.248638 –0.9721887 1.411789

1.301447 –0.6784622 –0.6184094 3.67147

Table 12.3 gives the Stata output for the random effect model using xtreg, re. This is the Swamy and Arora (1972) estimator which yields a price elasticity of -.420. This is closer to the Within estimator than the Between estimator. Table 12.3 Random Effects Estimator – Gasoline Demand Data Coef.

Std. Err.

T

P > |t|

log(Y /N ) log(PMG /PGDP ) log(Car/N ) Constant

0.5549858 –0.4203893 –0.6068402 1.996699

0.0591282 0.0399781 0.025515 0.184326

9.39 –10.52 –23.78 10.83

sigma u sigma e Rho

0.19554468 0.09233034 0.81769

(fraction of variance due to u i)

0.000 0.000 0.000 0.000

[95% Conf. Interval] 0.4390967 –0.498745 –0.6568487 1.635427

0.6708749 –0.3420336 –0.5568316 2.357971


12.4

Empirical Example

305

Table 12.4 Gasoline Demand Data. One-way Error Component Results

OLS WALHUS AMEMIYA SWAR IMLE ∗

β1 0.890 (0.036)∗ 0.545 (0.066) 0.602 (0.066) 0.555 (0.059) 0.588 (0.066)

β2 −0.892 (0.030)∗ −0.447 (0.046) −0.366 (0.042) −0.402 (0.042) −0.378 (0.046)

β3 −0.763 (0.019)∗ −0.605 (0.029) −0.621 (0.029) −0.607 (0.026) −0.616 (0.029)

ρ 0 0.75 0.93 0.82 0.91

These are biased standard errors when the true model has error component disturbances (see Moulton, 1986). Source: Baltagi and Griffin (1983). Reproduced by permission of Elsevier Science Publishers B.V. (North-Holland).

Table 12.5 Gasoline Demand Data. Wallace and Hussain (1969) Estimator Dependent Variable: GAS Method: Panel EGLS (Cross-section random effects) Sample: 1960 1978 Periods included: 19 Cross-sections included: 18 Total panel (balanced) observations: 342 Wallace and Hussain estimator of component variances

C log(Y /N ) log(PMG /PGDP ) log(Car/N )

Coefficient

Std. Error

t-Statistic

Prob.

1.938318 0.545202 −0.447490 −0.605086

0.201817 0.065555 0.045763 0.028838

9.604333 8.316682 −9.778438 −20.98191

0.0000 0.0000 0.0000 0.0000

S.D.

Rho

0.196715 0.113320

0.7508 0.2492

Effects Specification Cross-section random Idiosyncratic random

Table 12.4 gives the parameter estimates for OLS and three feasible GLS estimates of the slope coefficients along with their standard errors, and the corresponding estimate of ρ defined in (12.16). These were obtained using EViews by invoking the random effects estimation on the individual effects and choosing the estimation method from the options menu. Breusch’s (1987) iterative maximum likelihood was computed using Stata(xtreg, mle) and TSP. Table 12.5 gives the EViews output for the Wallace and Hussain (1969) random effects estimator, while Table 12.6 gives the EViews output for the Amemiya (1971) random effects estimator. Note that EViews calls the Amemiya estimator Wansbeek and Kapteyn (1989) since the latter paper generalizes this method to deal with unbalanced panels with missing observations, see Baltagi (2005) for details. Table 12.6 gives the Stata maximum likelihood output.


306

Chapter 12: Pooling Time-Series of Cross-Section Data

Table 12.6 Gasoline Demand Data. Wansbeek and Kapteyn (1989) Estimator Dependent Variable: GAS Method: Panel EGLS (Cross-section random effects) Sample: 1960 1978 Periods included: 19 Cross-sections included: 18 Total panel (balanced) observations: 342 Wallace and Hussain estimator of component variances

C log(Y /N ) log(PMG /PGDP ) log(Car/N )

Coefficient

Std. Error

t-Statistic

Prob.

2.188322 0.601969 −0.365500 −0.620725

0.216372 0.065876 0.041620 0.027356

10.11372 9.137941 −8.781832 −22.69053

0.0000 0.0000 0.0000 0.0000

S.D.

Rho

0.343826 0.092330

0.9327 0.0673

Effects Specification Cross-section random Idiosyncratic random

Table 12.7 Gasoline Demand Data. Random Effects Maximum Likelihood Estimator . xtreg c y p car,mle Random-effects ML regression Group variable (i): coun

Number of obs = Number of groups =

342 18

Random effects u i ∼ Gaussian

Obs per group: min = avg = max =

19 19.0 19

Log likelihood = 282.47697 c log(Y /N ) log(PMG /PGDP ) log(Car/N ) cons sigma u sigma e rho

LR chi2(3)

=

609.75

Prob > chi2

=

0.0000

Coef.

Std. Err.

z

P > |z|

.5881334 −.3780466 −.6163722 2.136168

.0659581 .0440663 .0272054 .2156039

8.92 −8.58 −22.66 9.91

0.000 0.000 0.000 0.000

.2922939 .0922537 .9094086

.0545496 .0036482 .0317608

[95% Conf. Interval] .4588578 −.464415 −.6696938 1.713593

.717409 −.2916782 −.5630506 2.558744

.2027512 .0853734 .8303747

.4213821 .0996885 .9571561

Likelihood-ratio test of sigma u = 0: chibar2(01)= 463.97 Prob >= chibar2 = 0.000


12.5

12.5

Testing in a Pooled Model

307

Testing in a Pooled Model

(1) The Chow-Test Before pooling the data one may be concerned whether the data is poolable. This hypothesis is also known as the stability of the regression equation across firms or across time. It can be formulated in terms of an unrestricted model which involves a separate regression equation for each firm yi = Zi δ i + ui

for i = 1, 2, . . . , N

(12.41)

where yi′ = (yi1 , . . . , yiT ), Zi = [ιT , Xi ] and Xi is (T × K). δ ′i is 1 × (K + 1) and ui is T × 1. The important thing to notice is that δ i is different for every regional equation. We want to test the hypothesis H0 ; δ i =δ for all i, versus H1 ; δ i = δ for some i. Under H0 we can write the restricted model given in (12.41) as: y = Zδ + u

(12.42)

′ ) where Z ′ = (Z1′ , Z2′ , . . . , ZN written as ⎛ Z1 0 . . . 0 ⎜ 0 Z2 . . . 0 ⎜ y=⎜ . .. .. ⎝ .. . .

0

0

. . . ZN

and u′ = (u′1 , u′2 , . . . , u′N ). The unrestricted model can also be ⎞⎛ ⎟⎜ ⎟⎜ ⎟⎜ ⎠⎝

δ1 δ2 .. . δN

⎟ ⎟ ⎟ + u = Z ∗ δ∗ + u ⎠

(12.43)

where δ ∗′ = (δ ′1 , δ ′2 , . . . , δ ′N ) and Z = Z ∗ I ∗ with I ∗ = (ιN ⊗ IK ′ ), an N K ′ × K ′ matrix, with K ′ = K + 1. Hence the variables in Z are all linear combinations of the variables in Z ∗ . Under the assumption that u ∼ N (0, σ 2 IN T ), the MVU estimator for δ in equation (12.42) is  δ OLS =  δM LE = (Z ′ Z)−1 Z ′ y

(12.44)

y = Z δ OLS + e

(12.45)

 δ i,OLS =  δ i,M LE = (Zi′ Zi )−1 Zi′ yi

(12.46)

yi = Zi δ i,OLS + ei

(12.47)

and therefore

implying that e = (IN T − Z(Z ′ Z)−1 Z ′ )y = M y = M (Zδ + u) = M u since MZ = 0. Similarly, under the alternative, the MVU for δ i is given by

and therefore

implying that ei = (IT − Zi (Zi′ Zi )−1 Zi′ )yi = Mi yi = Mi (Zi δ i + ui ) = Mi ui since Mi Zi = 0, and this is true for i = 1, 2, . . . , N . Also, let ⎞ ⎛ M1 0 . . . 0 ⎜ 0 M2 . . . 0 ⎟ ⎟ ⎜ M ∗ = IN T − Z ∗ (Z ∗′ Z ∗ )−1 Z ∗′ = ⎜ . .. ⎟ . .. ⎝ .. . ⎠ 0

0

. . . MN


308

Chapter 12: Pooling Time-Series of Cross-Section Data

∗ ∗ One can easily deduce that y = Z ∗ ˆδ + e∗ with e∗ = M ∗ y = M ∗ u and ˆδ = (Z ∗′ Z ∗ )−1 Z ∗′ y. Note that both M and M ∗ are symmetric and idempotent with M M ∗ = M ∗ . This easily follows since

Z(Z ′ Z)−1 Z ′ Z ∗ (Z ∗′ Z ∗ )−1 Z ∗′ = Z(Z ′ Z)−1 I ∗′ Z ∗′ Z ∗ (Z ∗′ Z ∗ )−1 Z ∗′ = Z(Z ′ Z)−1 Z ′ This uses the fact that Z = Z ∗ I ∗ . Now, e′ e − e∗′ e∗ = u′ (M − M ∗ )u and e∗′ e∗ = u′ M ∗ u are independent since (M − M ∗ )M ∗ = 0. Also, both quadratic forms when divided by σ 2 are distributed as χ2 since (M − M ∗ ) and M ∗ are idempotent, see Judge et al. (1985). Dividing these quadratic forms by their respective degrees of freedom, and taking their ratio leads to the following test statistic: Fobs = =

(e′ e − e∗′ e∗ )/(tr(M ) − tr(M ∗ )) e∗′ e∗ /tr(M ∗ )

(12.48)

(e′ e − e′1 e1 − e′2 e2 − .. − e′N eN )/(N − 1)K ′ (e′1 e1 + e′2 e2 + .. + e′N eN )/N (T − K ′ )

Under H0 , Fobs is distributed as an F ((N − 1)K ′ , N (T − K ′ )), see lemma 2.2 of Fisher (1970). This is exactly the Chow’s (1960) test extended to the case of N linear regressions. The URSS in this case is the sum of the N residual sum of squares obtained by applying OLS to (12.41), i.e., on each firm equation separately. The RRSS is simply the RSS from OLS performed on the pooled regression given by (12.42). In this case, there are (N −1)K ′ restrictions and the URSS has N (T − K ′ ) degrees of freedom. Similarly, one can test the stability of the regression across time. In this case, the degrees of freedom are (T − 1)K ′ and N (T − K ′ ) respectively. Both tests target the whole set of regression coefficients including the constant. If the LSDV model is suspected to be the proper specification, then the intercepts are allowed to vary but the slopes remain the same. To test the stability of the slopes only, the same Chowtest can be utilized, however the RRSS is now that of the LSDV regression with firm (or time) dummies only. The number of restrictions becomes (N − 1)K for testing the stability of the slopes across firms and (T − 1)K for testing their stability across time. The Chow-test however is proper under spherical disturbances, and if that hypothesis is not correct it will lead to improper inference. Baltagi (1981) showed that if the true specification of the disturbances is an error components structure then the Chow-test tend to reject poolability too often when in fact it is true. However, a generalization of the Chow-test which takes care of the general variance-covariance matrix is available in Zellner (1962). This is exactly the test of the null hypothesis H0 ; Rβ = r when Ω is that of the error components specification, see Chapter 9. Baltagi (1981) shows that this test performs well in Monte Carlo experiments. In this case, all we need to do is transform our model (under both the null and alternative hypotheses) such that the transformed disturbances have a variance of σ 2 IN T , then apply the Chow-test on the transformed model. The later step is legitimate because the transformed disturbances have homoskedastic variances and the usual Chow-test is legitimate. Given Ω = σ 2 Σ, we premultiply the restricted model given in (12.42) by Σ−1/2 and we call Σ−1/2 y = y, ˙ Σ−1/2 Z = Z˙ and −1/2 Σ u = u. ˙ Hence ˙ + u˙ y˙ = Zδ

(12.49)

with E(u˙ u˙ ′ ) = Σ−1/2 E(uu′ )Σ−1/2′ = σ 2 IN T . Similarly, we premultiply the unrestricted model given in (12.43) by Σ−1/2 and we call Σ−1/2 Z ∗ = Z˙ ∗ . Therefore


12.5

Testing in a Pooled Model

yË&#x2122; = ZË&#x2122; â&#x2C6;&#x2014; δ â&#x2C6;&#x2014; + uË&#x2122;

309

(12.50)

with E(uË&#x2122; uË&#x2122; â&#x20AC;˛ ) = Ď&#x192; 2 IN T . At this stage, we can test H0 ; δ i = δ for every i = 1, 2, . . . , N , simply by using the Chowstatistic, only now on the transformed models (12.49) and (12.50) since they satisfy uË&#x2122; â&#x2C6;ź N (0, Ď&#x192; 2 IN T ). Note that ZË&#x2122; = ZË&#x2122; â&#x2C6;&#x2014; I â&#x2C6;&#x2014; which is simply obtained from Z = Z â&#x2C6;&#x2014; I â&#x2C6;&#x2014; by premultiË&#x2122; ZË&#x2122; â&#x20AC;˛ Z) Ë&#x2122; â&#x2C6;&#x2019;1 ZË&#x2122; â&#x20AC;˛ , and MË&#x2122; â&#x2C6;&#x2014; = IN T â&#x2C6;&#x2019; ZË&#x2122; â&#x2C6;&#x2014; (ZË&#x2122; â&#x2C6;&#x2014;â&#x20AC;˛ ZË&#x2122; â&#x2C6;&#x2014; )â&#x2C6;&#x2019;1 ZË&#x2122; â&#x2C6;&#x2014;â&#x20AC;˛ , it is easy plying by ÎŁâ&#x2C6;&#x2019;1/2 . DeďŹ ning MË&#x2122; = IN T â&#x2C6;&#x2019; Z( â&#x2C6;&#x2014; Ë&#x2122; Ë&#x2122; to show that M and M are both symmetric, idempotent and such that MË&#x2122; MË&#x2122; â&#x2C6;&#x2014; = MË&#x2122; â&#x2C6;&#x2014; . Once again the conditions for lemma 2.2 of Fisher (1970) are satisďŹ ed, and the test-statistic (eË&#x2122;â&#x20AC;˛ eË&#x2122; â&#x2C6;&#x2019; eË&#x2122;â&#x2C6;&#x2014;â&#x20AC;˛ eË&#x2122;â&#x2C6;&#x2014; )/(tr(MË&#x2122; ) â&#x2C6;&#x2019; tr(MË&#x2122; â&#x2C6;&#x2014; )) â&#x2C6;ź F ((N â&#x2C6;&#x2019; 1)K â&#x20AC;˛ , N (T â&#x2C6;&#x2019; K â&#x20AC;˛ )) FË&#x2122;obs = eË&#x2122; â&#x2C6;&#x2014;â&#x20AC;˛ eË&#x2122; â&#x2C6;&#x2014; /tr(MË&#x2122; â&#x2C6;&#x2014; )

(12.51)

Ë&#x2122; â&#x2C6;&#x2019;1 ZË&#x2122; â&#x20AC;˛ yË&#x2122; implying that eË&#x2122; = MË&#x2122; yË&#x2122; = MË&#x2122; u. where eË&#x2122; = yË&#x2122; â&#x2C6;&#x2019; ZË&#x2122;  δË&#x2122; OLS and  δË&#x2122; OLS = (ZË&#x2122; â&#x20AC;˛ Z) Ë&#x2122; Similarly, â&#x2C6;&#x2014; â&#x2C6;&#x2014;   â&#x2C6;&#x2014;â&#x20AC;˛ â&#x2C6;&#x2014; â&#x2C6;&#x2019;1 â&#x2C6;&#x2014;â&#x20AC;˛ â&#x2C6;&#x2014; â&#x2C6;&#x2014; â&#x2C6;&#x2014; â&#x2C6;&#x2014; â&#x2C6;&#x2014; Ë&#x2122; This is the eË&#x2122; = yË&#x2122; â&#x2C6;&#x2019; ZË&#x2122; δË&#x2122; OLS and δË&#x2122; OLS = (ZË&#x2122; ZË&#x2122; ) ZË&#x2122; yË&#x2122; implying that eË&#x2122; = MË&#x2122; yË&#x2122; = MË&#x2122; u. Chow-test after premultiplying the model by ÎŁâ&#x2C6;&#x2019;1/2 or simply applying the Fuller and Battese (1974) transformation. See Baltagi (2005) for details. For the gasoline data in Baltagi and GriďŹ&#x192;n (1983), Chowâ&#x20AC;&#x2122;s test for poolability across countries yields an observed F -statistic of 129.38 and is distributed as F (68, 270) under H0 ; δ i = δ for i = 1, . . . , N . This tests the stability of four time-series regression coeďŹ&#x192;cients across 18 countries. The unrestricted SSE is based upon 18 OLS time-series regressions, one for each country. For the stability of the slope coeďŹ&#x192;cients only, H0 ; β i = β, an observed F -value of 27.33 is obtained which is distributed as F (51, 270) under the null. Chowâ&#x20AC;&#x2122;s test for poolability across time yields an F -value of 0.276 which is distributed as F (72, 266) under H0 ; δ t = δ for t = 1, . . . , T . This tests the stability of four cross-section regression coeďŹ&#x192;cients across 19 time periods. The unrestricted SSE is based upon 19 OLS cross-section regressions, one for each year. This does not reject poolability across time-periods. The test for poolability across countries, allowing for a one-way error components model yields an F -value of 21.64 which is distributed as F (68, 270) under H0 ; δ i = δ for i = 1, . . . , N . The test for poolability across time yields an F -value of 1.66 which is distributed as F (72, 266) under H0 ; δ t = δ for t = 1, . . . , T . This rejects H0 at the 5% level. (2) The Breusch-Pagan Test Next, we look at a Lagrange Multiplier test developed by Breusch and Pagan (1980), which tests whether H0 ; Ď&#x192; 2Îź = 0. The test statistic is given by 2  2 2 N T e ) â&#x2C6;&#x2019; 1 LM = (N T /2(T â&#x2C6;&#x2019; 1)) ( N e / t=1 it i=1 i. i=1

(12.52)

where eit denotes the OLS residuals on the pooled model, ei. denote their sum over t, respectively. Under the null hypothesis H0 this LM statistic is distributed as a Ď&#x2021;21 . For the gasoline data in Baltagi and GriďŹ&#x192;n (1983), the Breusch and Pagan LM test yields an LM statistic of 1465.6. This is obtained using the Stata command xtest0 after estimating the model with random effects. This is signiďŹ cant and rejects the null hypothesis. The corresponding likelihood ratio test assuming Normal disturbances is also reported by Stata maximum likelihood output for the random effects model. This yields an LR statistic of 463.97 which is asymptotically distributed as Ď&#x2021;21 under the null hypothesis H0 and is also signiďŹ cant.


310

Chapter 12: Pooling Time-Series of Cross-Section Data

One problem with the Breusch-Pagan test is that it assumes that the alternative hypothesis is two-sided when we know that σ 2μ > 0. A one-sided version of this test is given by Honda (1985): 4

′ e (IN ⊗ JT )e NT H0 − 1 → N (0, 1) (12.53) HO = 2(T − 1) e′ e where e denotes the vector of OLS residuals. Note that the square of this N (0, 1) statistic is the Breusch and Pagan (1980) LM test-statistic. Honda (1985) finds that this test statistic is uniformly most powerful and robust to non-normality. However, Moulton and Randolph (1989) showed that the asymptotic N (0, 1) approximation for this one-sided LM statistic can be poor even in large samples. They suggest an alternative Standardized Lagrange Multiplier (SLM) test whose asymptotic critical values are generally closer to the exact critical values than those of the LM test. This SLM test statistic centers and scales the one-sided LM statistic so that its mean is zero and its variance is one. SLM =

d − E(d) HO − E(HO)  =  var(HO) var(d)

(12.54)

where d = e′ De/e′ e and D = (IN ⊗ JT ). Using the results on moments of quadratic forms in regression residuals, see for e.g., Evans and King (1985), we get E(d) = tr(D P¯Z )/p and var(d) = 2{p tr(D P¯Z )2 − [tr(D P¯Z )]2 }/p2 (p + 2)

(12.55)

Z(Z ′ Z)−1 Z ′ .

Under the null hypothesis, SLM has an

where p = n − (K + 1) and P¯Z = In − asymptotic N (0, 1) distribution. (3) The Hausman-Test

A critical assumption in the error components regression model is that E(uit /Xit ) = 0. This is important given that the disturbances contain individual effects (the μi ’s) which are unobserved and may be correlated with the Xit ’s. For example, in an earnings equation these μi ’s may denote unobservable ability of the individual and this may be correlated with the schooling variable included on the right hand side of this equation. In this case, E(uit /Xit ) = 0 and the GLS  estimator β GLS becomes biased and inconsistent for β. However, the within transformation W ithin unbiased and consistent for β. wipes out these μi ’s and leaves the Within estimator β   Hausman (1978) suggests comparing β GLS and β W ithin , both of which are consistent under the null hypothesis H0 ; E(uit /Xit ) = 0, but which will have different probability limits if H0 is not   true. In fact, β W ithin is consistent whether H0 is true or not, while β GLS is BLUE, consistent and asymptotically efficient under H0 , but is inconsistent when H0 is false. A natural test statistic    would be based on q = β  = 0, and cov( q, β GLS − β W ithin . Under H0 , plim q GLS ) = 0. ′ Ω−1 X)−1 X ′ Ω−1 u and β   − β = (X − β = (X ′ QX)−1 X ′ Qu, Using the fact that β GLS W ithin one gets E( q ) = 0 and     cov(β , q) = var(β ) − cov(β ,β ) GLS

GLS

GLS

W ithin

= (X ′ Ω−1 X)−1 − (X ′ Ω−1 X)−1 X ′ Ω−1 E(uu′ )QX(X ′ QX)−1 = 0


12.6

Dynamic Panel Data Models

311

 W ithin = β  GLS − q, one gets Using the fact that β   var(β q ), W ithin ) = var(β GLS ) + var(

 since cov(β ) = 0. Therefore, GLS , q

2 ′ −1   var( q ) = var(β − (X ′ Ω−1 X)−1 W ithin ) − var(β GLS ) = σ ν (X QX)

(12.56)

q )]−1 q m = q′ [var(

(12.57)

 +w y ∗ = X ∗ β + Xγ

(12.58)

Hence, the Hausman test statistic is given by

and under H0 is asymptotically distributed as χ2K , where K denotes the dimension of slope  and vector β. In order to make this test operational, Ω is replaced by a consistent estimator Ω, GLS by its corresponding FGLS. An alternative asymptotically equivalent test can be obtained from the augmented regression

 = QX. Hausman’s test is now equivalent to where y ∗ = σ ν Ω−1/2 y, X ∗ = σ ν Ω−1/2 X and X  from testing whether γ = 0. This is a standard Wald test for the omission of the variables X (12.58). This test was generalized by Arellano (1993) to make it robust to heteroskedasticity and autocorrelation of arbitrary forms. In fact, if either heteroskedasticity or serial correlation is present, the variances of the Within and GLS estimators are not valid and the corresponding Hausman test statistic is inappropriate. For the Baltagi and Griffin (1983) gasoline data, the Hausman test statistic based on the difference between the Within estimator and that of feasible GLS based on Swamy and Arora (1972) yields a χ23 value of m = 306.1 which rejects the null hypothesis. This is obtained using the Stata command hausman.

12.6

Dynamic Panel Data Models

The dynamic error components regression is characterized by the presence of a lagged dependent variable among the regressors, i.e., yit = δyi,t−1 + x′it β + μi + ν it , i = 1, . . . , N ;

t = 1, . . . , T

(12.59)

where δ is a scalar, x′it is 1 × K and β is K × 1. This model has been extensively studied by Anderson and Hsiao (1982). Since yit is a function of μi , yi,t−1 is also a function of μi . Therefore, yi,t−1 , a right hand regressor in (12.59), is correlated with the error term. This renders the OLS estimator biased and inconsistent even if the ν it ’s are not serially correlated. For the FE estimator, the within transformation wipes out the μi ’s, but y˜i,t−1 will still be correlated with ν˜it even if the ν it ’s are not serially correlated. In fact, the Within estimator will be biased of O(1/T ) and its consistency will depend upon T being large, see Nickell (1981). An alternative transformation that wipes out the individual effects, yet does not create the above problem is the first difference (FD) transformation. In fact, Anderson and Hsiao (1982) suggested first differencing the model to get rid of the μi ’s and then using Δyi,t−2 = (yi,t−2 − yi,t−3 ) or simply yi,t−2 as an instrument for Δyi,t−1 = (yi,t−1 − yi,t−2 ). These instruments will not be


312

Chapter 12: Pooling Time-Series of Cross-Section Data

correlated with Δν it = ν i,t −ν i,t−1 , as long as the ν it ’s themselves are not serially correlated. This instrumental variable (IV) estimation method leads to consistent but not necessarily efficient estimates of the parameters in the model. This is because it does not make use of all the available moment conditions, see Ahn and Schmidt (1995), and it does not take into account the differenced structure on the residual disturbances (Δν it ). Arellano (1989) finds that for simple dynamic error components models the estimator that uses differences Δyi,t−2 rather than levels yi,t−2 for instruments has a singularity point and very large variances over a significant range of parameter values. In contrast, the estimator that uses instruments in levels, i.e., yi,t−2 , has no singularities and much smaller variances and is therefore recommended. Additional instruments can be obtained in a dynamic panel data model if one utilizes the orthogonality conditions that exist between lagged values of yit and the disturbances ν it . Let us illustrate this with the simple autoregressive model with no regressors: yit = δyi,t−1 + uit

i = 1, . . . , N

t = 1, . . . , T

(12.60)

where uit = μi + ν it with μi ∼ IID(0, σ 2μ ) and ν it ∼ IID(0, σ 2ν ), independent of each other and among themselves. In order to get a consistent estimate of δ as N → ∞ with T fixed, we first difference (12.60) to eliminate the individual effects yit − yi,t−1 = δ(yi,t−1 − yi,t−2 ) + (ν it − ν i,t−1 )

(12.61)

and note that (ν it − ν i,t−1 ) is MA(1) with unit root. For the first period we observe this relationship, i.e., t = 3, we have yi3 − yi2 = δ(yi2 − yi1 ) + (ν i3 − ν i2 ) In this case, yi1 is a valid instrument, since it is highly correlated with (yi2 − yi1 ) and not correlated with (ν i3 − ν i2 ) as long as the ν it are not serially correlated. But note what happens for t = 4, the second period we observe (12.61): yi4 − yi3 = δ(yi3 − yi2 ) + (ν i4 − ν i3 ) In this case, yi2 as well as yi1 are valid instruments for (yi3 − yi2 ), since both yi2 and yi1 are not correlated with (ν i4 − ν i3 ). One can continue in this fashion, adding an extra valid instrument with each forward period, so that for period T , the set of valid instruments becomes (yi1 , yi2 , . . . , yi,T −2 ). This instrumental variable procedure still does not account for the differenced error term in (12.61). In fact, E(Δν i Δν ′i ) = σ 2ν G where Δν ′i = (ν i3 − ν i2 , . . . , ν iT − ν i,T −1 ) and ⎛ 2 −1 0 ··· 0 0 0 ⎜ −1 2 −1 · · · 0 0 0 ⎜ ⎜ .. . . . . .. . .. .. .. .. .. G=⎜ . . ⎜ ⎝ 0 0 0 · · · −1 2 −1 0 0 0 ··· 0 −1 2

(12.62)

⎞ ⎟ ⎟ ⎟ ⎟ ⎟ ⎠


12.6

Dynamic Panel Data Models

is (T − 2) × (T − 2), since Δν i is MA(1) with unit root. Define ⎡ ⎤ [yi1 ] 0 ⎢ ⎥ [yi1 , yi2 ] ⎢ ⎥ Wi = ⎢ ⎥ . .. ⎣ ⎦ 0 [yi1 , . . . , yi,T −2 ]

313

(12.63)

Then, the matrix of instruments is W = [W1′ , . . . , WN′ ]′ and the moment equations described above are given by E(Wi′ Δν i ) = 0. Premultiplying the differenced equation (12.61) in vector form by W ′ , one gets W ′ Δy = W ′ (Δy−1 )δ + W ′ Δν

(12.64)

Performing GLS on (12.64) one gets the Arellano and Bond (1991) preliminary one-step consistent estimator  δ 1 = [(Δy−1 )′ W (W ′ (IN ⊗ G)W )−1 W ′ (Δy−1 )]−1 ′

−1

×[(Δy−1 ) W (W (IN ⊗ G)W )

(12.65)

W (Δy)]

The optimal generalized method of moments (GMM) estimator of δ 1 `a la Hansen (1982) for N → ∞ and T fixed using only the above moment restrictions yields the same expression as in (12.65) except that W ′ (IN ⊗ G)W =

N 5

Wi′ GWi

i=1

is replaced by VN =

N 5

Wi′ (Δν i )(Δν i )′ Wi

i=1

This GMM estimator requires no knowledge concerning the initial conditions or the distributions of ν i and μi . To operationalize this estimator, Δν is replaced by differenced residuals obtained from the preliminary consistent estimator  δ 1 . The resulting estimator is the two-step Arellano and Bond (1991) GMM estimator:  δ 2 = [(Δy−1 )′ W VN−1 W ′ (Δy−1 )]−1 [(Δy−1 )′ W VN−1 W ′ (Δy)]

(12.66)

 var( δ 2 ) = [(Δy−1 )′ W VN−1 W ′ (Δy−1 )]−1

(12.67)

A consistent estimate of the asymptotic var( δ 2 ) is given by the first term in (12.66),

δ 2 are asymptotically equivalent if the ν it are IID(0, σ 2ν ). Note that  δ 1 and  If there are additional strictly exogenous regressors xit as in (12.59) with E(xit ν is ) = 0 for all t, s = 1, 2, . . . , T, but where all the xit are correlated with μi , then all the xit are valid instruments for the first differenced equation of (12.59). Therefore, [x′i1 , x′i2 , . . . , x′iT ] should be added to each diagonal element of Wi in (12.63). In this case, (12.64) becomes W ′ Δy = W ′ (Δy−1 )δ + W ′ (ΔX)β + W ′ Δν


314

Chapter 12: Pooling Time-Series of Cross-Section Data

where ΔX is the stacked N (T − 2) × K matrix of observations on Δxit . One- and two-step estimators of (δ, β ′ )can be obtained from !  δ (12.68) = ([Δy−1 , ΔX]′ W VN−1 W ′ [Δy−1 , ΔX])−1 ([Δy−1 , ΔX]′ W VN−1 W ′ Δy)  β

as in (12.65) and (12.66). Arellano and Bond (1991) suggest Sargan’s (1958) test of over-identifying restrictions given by m = (Δ ν )′ W

N 5

Wi′ (Δ ν i )(Δ ν i )′ Wi

i=1

−1

W ′ (Δ ν ) ∼ χ2p−K−1

where p refers to the number of columns of W and Δ ν denote the residuals from a two-step estimation given in (12.68). To summarize, dynamic panel data estimation of equation (12.59) with individual fixed effects suffers from the Nickell (1981) bias. This disappears only if T tends to infinity. Alternatively, a GMM estimator was suggested by Arellano and Bond (1991) which basically differences the model to get rid of the individual specific effects and along with it any time invariant regressor. This also gets rid of any endogeneity that may be due to the correlation of these individual effects and the right hand side regressors. The moment conditions utilize the orthogonality conditions between the differenced errors and lagged values of the dependent variable. This assumes that the original disturbances are serially uncorrelated. In fact, two diagnostics are computed using the Arellano and Bond GMM procedure to test for first order and second order serial correlation in the disturbances. One should reject the null of the absence of first order serial correlation and not reject the absence of second order serial correlation. A special feature of dynamic panel data GMM estimation is that the number of moment conditions increase with T. Therefore, a Sargan test is performed to test the over-identification restrictions. There is convincing evidence that too many moment conditions introduce bias while increasing efficiency. It is even suggested that a subset of these moment conditions be used to take advantage of the trade-off between the reduction in bias and the loss in efficiency, see Baltagi (2005) for details. Arellano and Bond (1991) apply their GMM estimation and testing methods to a model of employment using a panel of 140 quoted UK companies for the period 1979-84. This is the benchmark data set used in Stata to obtain the one-step and two-step estimators described in (12.65) and (12.66) as well as the Sargan test for over-identification using the command (xtabond,twostep).The reader is asked to replicate their results in problem 22.

12.6.1

Empirical Illustration

Baltagi, Griffin and Xiong (2000) estimate a dynamic demand model for cigarettes based on panel data from 46 American states over 30 years 1963-1992. The estimated equation is ln Cit = α + β 1 ln Ci,t−1 + β 2 ln Pi,t + β 3 ln Yit + β 4 ln P nit + uit

(12.69)

where the subscript i denotes the ith state (i = 1, . . . , 46), and the subscript t denotes the tth year (t = 1, . . . , 30). Cit is real per capita sales of cigarettes by persons of smoking age (14


12.6

Dynamic Panel Data Models

315

years and older). This is measured in packs of cigarettes per head. Pit is the average retail price of a pack of cigarettes measured in real terms. Yit is real per capita disposable income. P nit denotes the minimum real price of cigarettes in any neighboring state. This last variable is a proxy for the casual smuggling effect across state borders. It acts as a substitute price attracting consumers from high-tax states like Massachusetts to cross over to New Hampshire where the tax is low. The disturbance term is specified as a two-way error component model: uit = μi + λt + ν it

i = 1, . . . , 46

t = 1, . . . , 30

(12.70)

where μi denotes a state-specific effect, and λt denotes a year-specific effect. The time-period effects (the λt ) are assumed fixed parameters to be estimated as coefficients of time dummies for each year in the sample. This can be justified given the numerous policy interventions as well as health warnings and Surgeon General’s reports. For example: (1) the imposition of warning labels by the Federal Trade Commission effective January 1965; (2) the application of the Fairness Doctrine Act to cigarette advertising in June 1967, which subsidized antismoking messages from 1968 to 1970; (3) the Congressional ban on broadcast advertising of cigarettes effective January 1971. The μi are state-specific effects which can represent any state-specific characteristic including the following: (1) States with Indian reservations like Montana, New Mexico and Arizona are among the biggest losers in tax revenues from non-Indians purchasing tax-exempt cigarettes from the reservations. (2) Florida, Texas, Washington and Georgia are among the biggest losers of revenues due to the purchasing of cigarettes from tax-exempt military bases in these states. (3) Utah, which has a high percentage of Mormon population (a religion which forbids smoking), has a per capita sales of cigarettes in 1988 of 55 packs, a little less than half the national average of 113 packs. (4) Nevada, which is a highly touristic state, has a per capita sales of cigarettes of 142 packs in 1988, 29 more packs than the national average. These state-specific effects may be assumed fixed, in which case one includes state dummy variables in equation (12.69). The resulting estimator is the Within estimator reported in Table 12.8. Comparing these estimates with OLS without state or time dummies, one can see that the coefficient of lagged consumption drops from 0.97 to 0.83 and the price elasticity goes up in absolute value from −0.09 to −0.30. The income elasticity switches sign from negative to positive going from −0.03 to 0.10. The OLS and Within estimators do not take into account the endogeneity of the lagged dependent variable, and therefore 2SLS and Within-2SLS are performed. The instruments used are one lag on price, neighboring price and income. These give lower estimates of lagged consumption and higher own price elasticities in absolute value. The Arellano and Bond (1991) two-step estimator yields an estimate of lagged consumption of 0.70 and a price elasticity of −0.40, both of which are significant. Sargan’s test for over-identification yields an observed value of 32.3. This is asymptotically distributed as χ227 and is not significant. This was obtained using the Stata command (xtabond2, twostep) with the collapse option to reduce the number of moment conditions used for estimation.


316

Chapter 12: Pooling Time-Series of Cross-Section Data

Table 12.8 Dynamic Demand for Cigarettes: 1963-92∗ OLS Within 2SLS Within-2SLS Arellano and Bond (two-step) ∗

lnCi,t−1 0.97 (157.7) 0.83 (66.3) 0.85 (25.3) 0.60 (17.0) 0.70 (10.2)

lnPit −0.090 (6.2) −0.299 (12.7) −0.205 (5.8) −0.496 (13.0) −0.396 (6.0)

lnYit −0.03 (5.1) 0.10 (4.2) −0.02 (2.2) 0.19 (6.4) 0.13 (3.5)

lnP nit 0.024 (1.8) 0.034 (1.2) 0.052 (3.1) −0.016 (0.5) −0.003 (0.1)

Numbers in parentheses are t-statistics. Source: Some of the results in this Table are reported in Baltagi, Griffin and Xiong (2000).

12.7

Program Evaluation and Difference-in-Differences Estimator

Suppose we want to study the effect of job training programs on earnings. An ideal experiment would assign individuals randomly, by a flip of a coin, to training and non-training camps, and then compare their earnings, holding other factors constant. This is a necessary experiment before the approval of any drug. Patients are randomly assigned to receive the drug or a placebo and the drug is approved or disapproved depending on the difference in the outcome between these two groups. In this case, the FDA is concerned with the drug’s safety and its effectiveness. However, we run into problems in setting this experiment. How can we hold other factors constant? Even twins which have been used in economic studies are not identical and may have different life experiences. The individual’s prior work experience will affect one’s chances in getting a job after training. But as long as the individuals are randomly assigned, the distribution of work experience is the same in the treatment and control group, i.e., participation in the job training is independent of prior work experience. In this case, omitting previous work experience from the analysis will not cause omitted variable bias in the estimator of the effect of the training program on future employment. Stock and Watson (2003) discuss threats to the internal and external validity of such experiments. The former include: (i) failure to randomize, or (ii) to follow the treatment protocol. These failures can cause bias in estimating the effect of the treatment. The first can happen when individuals are assigned non-randomly to the treatment and non-treatment groups. The second can happen, for example, when some people in the training program do not show up for all training sessions; or when some people who are not supposed to be in the training program are allowed to attend some of these training sessions. Attrition caused by people dropping out of the experiment in either group can cause bias especially if the cause of attrition is related to their acquiring or not acquiring training. In addition, small samples, usually associated with expensive experiments, can affect the precision of the estimates. There can also be experimental effects, brought about by people trying harder simply because the worker being trained feels noticed or because the trainer has a stake in the success of the program. Stock and Watson (2003, p. 380) argue that “threats to external validity compromise the ability to generalize the results of the experiment to other populations and settings. Two such threats are when the experimental sample is not representative of the population of interest


Problems

317

and when the treatment being studied is not representative of the treatment that would be implemented more broadly.” They also warn about “general equilibrium effects”where, for example, turning a small, temporary experimental program into a widespread, permanent program might change the economic environment sufficiently that the results of the experiment cannot be generalized. For example, it could displace employer-provided training, thereby reducing the net benefits of the program. 12.7.1

The Difference-in-Differences Estimator

With panel data, observations on the same subjects before and after the training program allow us to estimate the effect of this program on earnings. In simple regression form, assuming the assignment to the training program is random, one regresses the change in earnings before and after training is completed on a dummy variable which takes the value 1 if the individual received training and zero if they did not. This regression computes the average change in earnings for the treatment group before and after the training program and subtracts that from the average change in earnings for the control group. One can include additional regressors which measure the individual characteristics prior to training. Examples are gender, race, education and age of the individual. Card (1990) used a quasi-experiment to see whether immigration reduces wages. Taking advantage of the “Mariel boatlift” where a large number of Cuban immigrants entered Miami. Card (1990) used the difference-in-differences estimator, comparing the change in wages of lowskilled workers in Miami to the change in wages of similar workers in other comparable U.S. cities over the same period. Card concluded that the influx of Cuban immigrants had a negligible effect on wages of less-skilled workers.

Problems 1.

(a) Premultiply (12.11) by Q and verify that the transformed equation reduces to (12.12). Show that the new disturbances Qν have zero mean and variance-covariance matrix σ 2ν Q. Hint: QZμ = 0. (b) Show that the GLS estimator is the same as the OLS estimator on this transformed regression equation. Hint: Use one of the necessary and sufficient conditions for GLS to be equivalent to OLS given in Chapter 9.

2.

(c) Using the Frisch-Waugh-Lovell Theorem given in Chapter 7, show that the estimator derived  = (X ′ QX)−1 X ′ Qy. in part (b) is the Within estimator and is given by β

(a) Show that Ω given in (12.17) can be written as (12.18).

(b) Show that P and Q are symmetric, idempotent, orthogonal and sum to the identity matrix. (c) For Ω−1 given by (12.19), verify that ΩΩ−1 = Ω−1 Ω = IN T . (d) For Ω−1/2 given by (12.20), verify that Ω−1/2 Ω−1/2 = Ω−1 . 3. Premultiply y by σ ν Ω−1/2 where Ω−1/2 is defined in (12.20) and show that the resulting y ∗ has a ∗ = yit − θ¯ yi. , where the θ = 1 − σ ν /σ 1 and σ 21 = T σ2μ + σ 2ν . typical element yit σ 2ν ) = σ 2ν . Hint: E(u′ Qu) = 4. Using (12.21) and (12.22), show that E( σ 21 ) = σ 21 and E( ′ ′ ′ E{tr(u Qu)} = E{tr(uu Q)} = tr{E(uu )Q} = tr(ΩQ).


318 5.

6.

Chapter 12: Pooling Time-Series of Cross-Section Data 2  (a) Show that σ  ν given in (12.23) is unbiased for σ 2ν .

2  (b) Show that σ  1 given in (12.26) is unbiased for σ 21 .

(a) Perform OLS on the system of equations given in (12.27) and show that the resulting estimator is  δ OLS = (Z ′ Z)−1 Z ′ y. (b) Perform GLS on this system of equations and show that the resulting estimator is  δ GLS = (Z ′ Ω−1 Z)−1 Z ′ Ω−1 y where Ω−1 is given in (12.19).

−1 2   7. Using the var(β GLS ) expression below (12.30) and var(β W ithin ) = σ ν WXX , show that −1 −1   (var(β − (var(β = φ2 BXX /σ 2ν GLS )) W ithin ))

8.

  which is positive semi-definite. Conclude that var(β W ithin )− var(β GLS ) is positive semi-definite.

(a) Using the concentrated likelihood function in (12.34), solve ∂Lc /∂φ2 = 0 and verify (12.35).

(b) Solve ∂Lc /∂β = 0 and verify (12.36). 9.

(a) For the predictor of yi,T +S given in (12.37), compute E(ui,T +S uit ) for t = 1, 2, . . . , T and verify that w = E(ui,T +S u) = σ 2μ (ℓi ⊗ ιT ) where ℓi is the i-th column of IN .

(b) Verify (12.39) by showing that (ℓ′i ⊗ ι′T )P = (ℓ′i ⊗ ι′T ).

10. Using the gasoline demand data of Baltagi and Griffin (1983), given on the Springer web site as GASOLINE.DAT, reproduce Tables 12.1 through 12.7. 11. For the random one-way error components model given in (12.1) and (12.2), consider the OLS estimator of var(uit ) = σ 2 , which is given by s2 = e′ e/(n − K ′ ), where e denotes the vector of OLS residuals, n = N T and K ′ = K + 1. (a) Show that E(s2 ) = σ 2 + σ 2μ [K ′ − tr(IN ⊗ JT )PX ]/(n − K ′ ).

(b) Consider the inequalities given by Kiviet and Kr¨ amer (1992) which state that 0 ≤ mean of n − K ′ smallest roots of Ω ≤ E(s2 ) ≤ mean of n − K ′ largest roots of Ω ≤ tr(Ω)/(n − K ′ ) where Ω = E(uu′ ). Show that for the one-way error components model, these bounds are 0 ≤ σ 2ν + (n − T K ′)σ 2μ /(n − K ′ ) ≤ E(s2 ) ≤ σ 2ν + nσ 2μ /(n − K ′ ) ≤ nσ 2 /(n − K ′ ). As n → ∞, both bounds tend to σ 2 , and s2 is asymptotically unbiased, irrespective of the particular evolution of X, see Baltagi and Kr¨ amer (1994) for a proof of this result. 12. Verify the relationship between M and M ∗ , i.e., M M ∗ = M ∗ , given below (12.47). Hint: Use the fact that Z = Z ∗ I ∗ with I ∗ = (ιN ⊗ IK ′ ). 13. Verify that M˙ and M˙ ∗ defined below (12.50) are both symmetric, idempotent and satisfy M˙ M˙ ∗ = M˙ ∗ . 14. For the gasoline data used in problem 10, verify the Chow-test results given below equation (12.51). 15. For the gasoline data, compute the Breusch-Pagan, Honda and Standardized LM tests for H0 ; σ 2μ = 0.  denotes the LSDV estimator and β  16. If β GLS denotes the GLS estimator, then    (a) Show that q = β q, β GLS − β satisfies cov( GLS ) = 0.

(b) Verify equation (12.56).


Problems

319

17. For the gasoline data used in problem 10, replicate the Hausman test results given below equation (12.58). 18. For the cigarette data given as CIGAR.TXT on the Springer web site, reproduce the results given in Table 12.8. See also Baltagi, Griffin and Xiong (2000). 19. Heteroskedastic Fixed Effects Models. This is based on Baltagi (1996). Consider the fixed effects model yit = αi + uit

i = 1, 2, . . . , N ; t = 1, 2, . . . , Ti

where yit denotes output in industry i at time t and αi denotes the industry fixed effect. The disturbances uit are assumed to be independent with heteroskedastic variances σ 2i . Note that the data are unbalanced with different number of observations for each industry. (a) Show that OLS and GLS estimates of αi are identical. N N (b) Let σ 2 = i=1 Ti σ 2i /n where n = i=1 Ti , be the average disturbance variance. Show that the GLS estimator of σ 2 is unbiased, whereas the OLS estimator of σ 2 is biased. Also show that this bias disappears if the data are balanced or the variances are homoskedastic. (c) Define λ2i = σ 2i /σ 2 for i = 1, 2 . . . , N . Show that for α′ = (α1 , α2 , . . . , αN ) E[estimated var( αOLS ) − true var( αOLS )] = σ 2 [(n −

N 5 i=1

λ2i )/(n − N )] diag (1/Ti ) − σ 2 diag (λ2i /Ti )

This problem shows that in case there are no regressors in the unbalanced panel data model, fixed effects with heteroskedastic disturbances can be estimated by OLS, but one has to correct the standard errors. 20. The Relative Efficiency of the Between Estimator with Respect to the Within Estimator. This is based on Baltagi (1999). Consider the simple panel data regression model yit = α + βxit + uit

i = 1, 2, . . . , N ;

t = 1, 2, . . . , T

(1)

where α and β are scalars. Subtract the mean equation to get rid of the constant ¯.. ) + uit − u¯.. , yit − y¯.. = β(xit − x

(2)

T where x ¯.. = ΣN ¯.. and u¯.. are similarly defined. Add and subtract x¯i. from the i=1 Σt=1 xit /N T and y regressor in parentheses and rearrange

yit − y¯.. = β(xit − x ¯i. ) + β(¯ xi. − x ¯.. ) + uit − u ¯..

(3)

where x¯i. = ΣTt=1 xit /T . Now run the unrestricted least squares regression yit − y¯.. = β w (xit − x¯i. ) + β b (¯ xi. − x¯.. ) + uit − u ¯.. where β w is not necessarily equal to β b .

(4)


320

Chapter 12: Pooling Time-Series of Cross-Section Data (a) Show that the least squares estimator of β w from (4) is the Within estimator and that of β b is the Between estimator. (b) Show that if uit = μi + ν it where μi ∼ IID(0, σ 2μ ) and ν it ∼ IID(0, σ 2ν ) independent of each other and among themselves, then ordinary least squares (OLS) is equivalent to generalized least squares (GLS) on (4). (c) Show that for model (1), the relative efficiency of the Between estimator with respect to the Within estimator is equal to (BXX /WXX )[(1 − ρ)/(T ρ + (1 − ρ))], where WXX = T ΣN ¯i. )2 denotes the Within variation and BXX = T ΣN xi. − x ¯.. )2 denotes i=1 Σt=1 (xit − x i=1 (¯ 2 2 2 the Between variation. Also, ρ = σ μ /(σ μ + σ ν ) denotes the equicorrelation coefficient. (d) Show that the square of the t-statistic used to test H0 ; β w = β b in (4) yields exactly Hausman’s (1978) specification test.

21. For the crime example of Cornwell and Trumbull (1994) studied in Chapter 11. Use the panel data given as CRIME.DAT on the Springer web site to replicate the Between and Within estimates given in Table 1 of Cornwell and Trumbull (1994). Compute 2SLS and Within-2SLS (2SLS with countu dummies) using offense mix and per capita tax revenue as instruments for the probability of arrest and police per capita. Comment on the results. 22. Consider the Arellano and Bond (1991) dynamic employment equation for 140 UK companies over the period 1979-1984. Replicate all the estimation results in Table 4 of Arellano and Bond (1991, p. 290).

References This chapter is based on Baltagi (2005). Ahn, S.C. and P. Schmidt (1995), “Efficient Estimation of Models for Dynamic Panel Data,” Journal of Econometrics, 68: 5-27. Amemiya, T. (1971), “The Estimation of the Variances in a Variance-Components Model,” International Economic Review, 12: 1-13. Anderson, T.W. and C. Hsiao (1982), “Formulation and Estimation of Dynamic Models Using Panel Data, Journal of Econometrics, 18: 47-82. Arellano, M. (1989), “A Note on the Anderson-Hsiao Estimator for Panel Data,” Economics Letters, 31: 337-341. Arellano, M. (1993), “On the Testing of Correlated Effects With Panel Data,” Journal of Econometrics, 59: 87-97. Arellano, M. and S. Bond (1991), “Some Tests of Specification for Panel Data: Monte Carlo Evidence and An Application to Employment Equations,” Review of Economic Studies, 58: 277-297. Balestra, P. (1973), “Best Quadratic Unbiased Estimators of the Variance-Covariance Matrix in Normal Regression,” Journal of Econometrics, 2: 17-28. Baltagi, B.H. (1981), “Pooling: An Experimental Study of Alternative Testing and Estimation Procedures in a Two-Way Errors Components Model,” Journal of Econometrics, 17: 21-49. Baltagi, B.H. (1996), “Heteroskedastic Fixed Effects Models,” Problem 96.5.1, Econometric Theory, 12: 867.


References

321

Baltagi, B.H. (1999), “The Relative Efficiency of the Between Estimator with Respect to the Within Estimator,” Problem 99.4.3, Econometric Theory, 15: 630-631. Baltagi, B.H. (2005), Econometric Analysis of Panel Data (Wiley: Chichester). Baltagi, B.H. and J.M. Griffin (1983), “Gasoline Demand in the OECD: An Application of Pooling and Testing Procedures,” European Economic Review, 22: 117-137. Baltagi, B.H., J.M. Griffin and W. Xiong (2000), “To Pool or Not to Pool: Homogeneous Versus Heterogeneous Estimators Applied to Cigarette Demand,” Review of Economics and Statistics, 82: 117-126. Baltagi, B.H. and W. Kr¨ amer (1994), “Consistency, Asymptotic Unbiasedness and Bounds on the Bias of s2 in the Linear Regression Model with Error Components Disturbances,” Statistical Papers, 35: 323-328. Breusch, T.S. (1987), “Maximum Likelihood Estimation of Random Effects Models,” Journal of Econometrics, 36: 383-389. Breusch, T.S. and A.R. Pagan (1980), “The Lagrange Multiplier Test and its Applications to Model Specification in Econometrics,” Review of Economic Studies, 47: 239-253. Card (1990), “The Impact of the Mariel Boat Lift on the Miami Labor Market,” Industrial and Labor Relations Review, 43: 245-253. Chow, G.C. (1960), “Tests of Equality Between Sets of Coefficients in Two Linear Regressions,” Econometrica, 28: 591-605. Cornwell, C. and W.N. Trumbull (1994), “Estimating the Economic Model of Crime with Panel Data,” Review of Economics and Statistics 76: 360-366. Evans, M.A. and M.L. King (1985), “Critical Value Approximations for Tests of Linear Regression Disturbances,” Australian Journal of Statistics, 27: 68-83. Fisher, F.M. (1970), “Tests of Equality Between Sets of Coefficients in Two Linear Regressions: An Expository Note,” Econometrica, 38: 361-366. Fuller, W.A. and G.E. Battese (1974), “Estimation of Linear Models with Cross-Error Structure,” Journal of Econometrics, 2: 67-78. Goldberger, A.S. (1962), “Best Linear Unbiased Prediction in the Generalized Linear Regression Model,” Journal of the American Statistical Association, 57: 369-375. Graybill, F.A. (1961), An Introduction to Linear Statistical Models (McGraw-Hill: New York). Hansen, L.P. (1982), “Large Sample Properties of Generalized Method of Moments Estimators,” Econometrica, 50: 1029-1054. Hausman, J.A. (1978), “Specification Tests in Econometrics,” Econometrica, 46: 1251-1271. Honda, Y. (1985), “Testing the Error Components Model with Non-Normal Disturbances,” Review of Economic Studies, 52: 681-690. Hsiao, C. (2003), Analysis of Panel Data (Cambridge University Press: Cambridge). Judge, G.G., W.E. Griffiths, R.C. Hill, H. Lutkepohl and T.C. Lee (1985), The Theory and Practice of Econometrics (Wiley: New York). Kiviet, J.F. and W. Kr¨ amer (1992), “Bias of s2 in the Linear Regression Model with Correlated Errors,” Empirical Economics, 16: 375-377.


322

Chapter 12: Pooling Time-Series of Cross-Section Data

Maddala, G.S. (1971), “The Use of Variance Components Models in Pooling Cross Section and Time Series Data,” Econometrica, 39: 341-358. Maddala, G.S. and T. Mount (1973), “A Comparative Study of Alternative Estimators for Variance Components Models Used in Econometric Applications,” Journal of the American Statistical Association, 68: 324-328. Moulton, B.R. and W.C. Randolph (1989), “Alternative Tests of the Error Components Model,” Econometrica, 57: 685-693. Nerlove, M. (1971), “A Note on Error Components Models,” Econometrica, 39: 383-396. Nickell, S. (1981), “Biases in Dynamic Models with Fixed Effects,”Econometrica, 49: 1417-1426. Searle, S.R. (1971), Linear Models (Wiley: New York). Sargan, J. (1958), “The Estimation of Economic Relationships Using Instrumental Variables,” Econometrica, 26: 393-415. Swamy, P.A.V.B. and S.S. Arora (1972), “The Exact Finite Sample Properties of the Estimators of Coefficients in the Error Components Regression Models,” Econometrica, 40: 261-275. Taub, A.J. (1979), “Prediction in the Context of the Variance-Components Model,” Journal of Econometrics, 10: 103-108. Taylor, W.E. (1980), “Small Sample Considerations in Estimation from Panel Data,” Journal of Econometrics, 13: 203-223. Wallace, T. and A. Hussain (1969), “The Use of Error Components Models in Combining Cross-Section and Time-Series Data,” Econometrica, 37: 55-72. Wansbeek, T.J. and A. Kapteyn (1978), “The Separation of Individual Variation and Systematic Change in the Analysis of Panel Data,” Annales de l’INSEE, 30-31: 659-680. Wansbeek, T.J. and A. Kapteyn (1982), “A Simple Way to Obtain the Spectral Decomposition of Variance Components Models for Balanced Data,” Communications in Statistics All, 2105-2112. Wansbeek, T.J. and A. Kapteyn, (1989), “Estimation of the error components model with incomplete panels,” Journal of Econometrics 41: 341–361. Zellner, A. (1962), “An Efficient Method of Estimating Seemingly Unrelated Regression and Tests for Aggregation Bias,” Journal of the American Statistical Association, 57: 348-368.


CHAPTER 13

Limited Dependent Variables 13.1

Introduction

In labor economics, one is faced with explaining the decision to participate in the labor force, the decision to join a union, or the decision to migrate from one region to the other. In finance, a consumer defaults on a loan or a credit card debt, or purchases a stock or an asset like a house or a car. In these examples, the dependent variable is usually a dummy variable with values 1 if the worker participates (or consumer defaults on a loan) and 0 if he or she does not participate (or default). We dealt with dummy variables as explanatory variables on the right hand side of the regression, but what additional problems arise when this dummy variable appears on the left hand side of the equation? As we have done in previous chapters, we first study its effects on the usual least squares estimator, and then consider alternative estimators that are more appropriate for models of this nature.

13.2

The Linear Probability Model

What is wrong with running OLS on this model? After all, it is a feasible procedure. For the labor force participation example one regresses the dummy variable for participation on age, sex, race, marital status, number of children, experience and education, etc. The prediction from this OLS regression is interpreted as the likelihood of participating in the labor force. The problems with this interpretation are the following: (i) We are predicting probabilities of participation for each individual, whereas the actual values observed are 0 or 1. (ii) There is no guarantee that yi , the predicted value of yi is going to be between 0 and 1. In fact, one can always find values of the explanatory variables that would generate a corresponding prediction outside the (0, 1) range. (iii) Even if one is willing to assume that the true model is a linear regression given by yi = x′i β + ui

i = 1, 2, . . . , n.

(13.1)

what properties does this entail on the disturbances? It is obvious that yi = 1 only when ui = 1 − x′i β, let us say with probability π i , where π i is to be determined. Then yi = 0 only when ui = −x′i β with probability (1 − π i ). For the disturbances to have zero mean E(ui ) = π i (1 − x′i β) + (1 − π i )(−x′i β) = 0

(13.2)

Solving for π i , one gets that π i = x′i β. This also means that var(ui ) = π i (1 − π i ) = x′i β(1 − x′i β)

(13.3)

which is heteroskedastic. Goldberger (1964) suggests correcting for this heteroskedasticity by ′  first running OLS to estimate β, and estimating σ 2i = var(ui ) by σ 2i = x′i β OLS (1 − xi β OLS ) =


324

Chapter 13: Limited Dependent Variables

yi (1 − yi ). In the next step a Weighted Least Squares (WLS) procedure is run on (13.1) with the original observations divided by σ i . One cannot compute σ i if OLS predicts yi larger than 1 or smaller than 0. Suggestions in the literature include substituting 0.005 instead of yi < 0, and 0.995 for yi > 1. However, these procedures do not perform well, and the WLS predictions themselves are not guaranteed to fall in the (0, 1) range. Therefore, one should use the robust White heteroskedastic variance-covariance matrix option when estimating linear probability models, otherwise the standard errors are biased and inference is misleading. This brings us to the fundamental problem with OLS, i.e., its functional form. We are trying to predict yi = F (x′i β) + ui

(13.4)

with a linear regression equation, see Figure 13.1, where the more reasonable functional form for this probability is an S-shaped cumulative distribution functional form. This was justified in the biometrics literature as follows: An insect has a tolerance to an insecticide Ii∗ , which is an unobserved random variable with cumulative distribution function (c.d.f.) F . If the dosage of insecticide administered induces a stimulus Ii that exceeds Ii∗ , the insect dies, i.e., yi = 1. Therefore Pr(yi = 1) = Pr(Ii∗ ≤ Ii ) = F (Ii )

(13.5)

To put it in an economic context, Ii∗ could be the unobserved reservation wage of a worker, and if we increase the offered wage beyond that reservation wage, the worker participates in the labor force. In general, Ii could be represented as a function of the individuals characteristics, i.e., the xi ’s. F (x′i β) is by definition between zero and 1 for all values of xi . Also, the linear probability model yields the result that ∂π i /∂xk = β k , for every i. This means that the probability of participating (π i ) always changes at the same rate with respect to unit increases in the offer wage xk . However, this probability model gives ∂π i /∂xk = [∂F (zi )/∂zi ] · [∂zi /∂xk ] = f (x′i β) · β k

(13.6)

where zi = x′i β, and f is the probability density function (p.d.f.). Equation (13.6) makes more sense because if xk denotes the offered wage, changing the probability of participation π i from 0.96 to 0.97 requires a larger change in xk than changing π i from 0.23 to 0.24. If F (x′i β) is the true probability function, assuming it is linear introduces misspecification, and as Figure 13.1 indicates, for xi < xℓ , all the ui ’s generated by a linear probability approximation are positive. Similarly for all xi > xu , all the ui ’s generated by a linear probability approximation are negative.

13.3

Functional Form: Logit and Probit

Having pointed out the problems with considering the functional form F as linear, we turn to two popular functional forms of F , the logit and the probit. These two c.d.f.’s differ only in the tails, and the logit resembles the c.d.f. of a t-distribution with 7 degrees of freedom, whereas the probit is the normal c.d.f., or that of a t with ∞ degrees of freedom. Therefore, these two forms will give similar predictions unless there are an extreme number of observations in the tails.


13.3 Pr(O)

Functional Form: Logit and Probit

325

= + >N –

1

6HKA 2HO / N) +

0

+

N

NK

x

Figure 13.1 Linear Probability Model

√ z 2 We will use the conventional notation Φ(z) = −∞ φ(u)du, where φ(z) = e−z /2 / 2π for −∞ < z < ∞, for the probit. Also, Λ(z) = ez /(1 + ez ) = 1/(1 + e−z ) for −∞ < z < +∞, for the logit. Some results that we will use quite often in our derivations are the following: dΦ/dz = φ, and dΛ/dz = Λ(1 − Λ). The p.d.f. of the logistic distribution is the product of its c.d.f. and one minus this c.d.f. Therefore, the marginal effects considered above for a general F are respectively, ∂Φ(x′i β)/∂xk = φi β k

(13.7)

∂Λ(x′i β)/∂xk = Λi (1 − Λi )β k

(13.8)

and

where φi = φ(x′i β) and Λi = Λ(x′i β). One has to be careful with the computation of partial derivatives in case there is a dummy variable among the explanatory variables. For such models, one should compute the marginal effects of a change in one unit of a continuous variable xk for both values of the dummy variable.

Illustrative Example: Using the probit model, suppose that the probability of joining a union is estimated as follows: π i = Φ(2.5 − 0.06 WKS i + 0.95 OCC i ) where WKS is the number of weeks worked and OCC = 1, if the individual is in a blue-collar occupation, and zero otherwise. Weeks worked in this sample range from 20 to 50. From (13.7), the marginal effect of one extra week of work on the probability of joining the union is given by:


326

Chapter 13: Limited Dependent Variables

For Blue-Collar Workers:

−0.06 φ[2.5 − 0.06 WKS + 0.95] = −0.06 φ[2.25] = −0.002 at WKS = 20 = −0.06 φ[1.35] = −0.010 at WKS = 35 = −0.06 φ[0.45] = −0.022 at WKS = 50

For Non Blue-Collar Workers:

−0.06 φ[2.5 − 0.06 WKS ] = −0.06 φ[1.3] = −0.010 at WKS = 20 = −0.06 φ[0.4] = −0.022 at WKS = 35 = −0.06 φ[−0.5] = −0.021 at WKS = 50

Note how different these marginal effects are for blue-collar versus non blue-collar workers even for the same weeks worked. Increasing weeks worked from 20 to 21 reduces the probability of joining the union by 0.002 for a Blue-Collar worker. This is compared to five times that amount for a Non Blue-Collar worker.

13.4

Grouped Data

In the biometrics literature, grouped data is very likely from laboratory experiments, see Cox (1970). In the insecticide example, every dosage level xi is administered to a group of insects of size ni , and the proportion of insects that die are recorded (pi ). This is done for i = 1, 2, . . . , M dosage levels. P [yi = 1] = π i = P [Ii∗ ≤ Ii ] = Φ(α + βxi ) where Ii∗ is the tolerance and Ii = α + βxi is the stimulus. In economics, observations may be grouped by income levels or age and we observe the labor participation rate for each income or age group. For this type of grouped data, we estimate the probability of participating in the labor force π i with pi , the proportion from the sample. This requires a large number of observations in each group, i.e., a large ni for i = 1, 2, . . . , M . In this case, the approximation is zi = Φ−1 (pi ) ∼ = α + βxi

(13.9)

for each pi , we compute the standardized normal variates, the zi ’s, and we have an estimate of α + βxi . Note that the standard normal distribution assumption is not restrictive in the sense that if Ii∗ is N (μ, σ 2 ) rather than N (0, 1), then one standardizes the P [Ii∗ ≤ Ii ] by subtracting μ and dividing by σ, in which case the new Ii∗ is N (0, 1) and the new α is (α − μ)/σ, whereas the new β is β/σ. This also implies that μ and σ are not separately estimable. A plot of the zi ’s versus the xi ’s would give estimates of α and β. For the biometrics example, one can compute LD50, which is the dosage level that will kill 50% of the insect population. This corresponds to  Similarly, LD95 corresponds to zi = 1.645, which solves for zi = 0, which solves for xi = − α/β.  Alternatively, for the economic example, LD50 is the minimum reservation xi = (1.645 − α  )/β. wage that is necessary for a 50% labor participation rate. One could improve on this method by including more x’s on the right hand side of (13.9). In this case, one can no longer plot the zi values versus the x variables. However, one can run


13.4

Grouped Data

327

OLS of z on these x’s. One problem remains, OLS ignores the heteroskedasticity in the error term. To see this: pi = π i + ǫi = F (x′i β) + ǫi

(13.10)

where F is a general c.d.f. and π i = F (x′i β). Using the properties of the binomial distribution, E(pi ) = π i and var(pi ) = π i (1 − π i )/ni . Defining zi = F −1 (pi ), we obtain from (13.10) zi = F −1 (pi ) = F −1 (π i + ǫi ) ∼ = F −1 (π i ) + [dF −1 (π i )/dπ i ]ǫi

(13.11)

where the approximation ∼ = is a Taylor series expansion around ǫi = 0. Since F is monotonic π i = F (F −1 (π i )). Let wi = F −1 (π i ) = x′i β, differentiating with respect to π gives 1 = [dF (wi )/dwi ]dwi /dπ i

(13.12)

Alternatively, this can be rewritten as dF −1 (π i )/dπ i = dwi /dπ i = 1/{dF (wi )/dwi } = 1/f (wi ) = 1/f (x′i β)

(13.13)

where f is the probability density function corresponding to F . Using (13.13), equation (13.11) can be rewritten as zi = F −1 (pi ) ∼ = F −1 (π i ) + ǫi /f (x′i β) = F

−1

(F (x′i β))

+

ǫi /f (x′i β)

=

x′i β

(13.14) + ǫi /f (x′i β)

From (13.14), it is clear that the regression disturbances of zi on xi are given by ui ∼ = ǫi /f (x′i β), 2 2 ′ 2 with E(ui ) = 0 and σ i = var(ui ) = var(ǫi )/f (xi β) = π i (1 − π i )/(ni fi ) = Fi (1 − Fi )/(ni fi2 ) since π i = Fi where the subscript i on f or F denotes that the argument of that function is x′i β . This heteroskedasticity in the disturbances renders OLS on (13.14) consistent but inefficient. For the probit, σ 2i = Φi (1 − Φi )/(ni φ2i ), and for the logit, σ 2i = 1/[ni Λi (1 − Λi )], since fi = Λi (1 − Λi ). Using 1/σ i as weights, a WLS procedure can be performed on (13.14). Note that F −1 (p) for the logit is simply log[p/(1 − p)]. This is one more reason why the logistic functional form is so popular. In this case one regresses log[p/(1 − p)] on x correcting for heteroskedasticity using WLS. This procedure is also known as the minimum logit chi-square method and is due to Berkson (1953). In order to obtain feasible estimates of the σ i ’s, one could use the OLS estimates of β from (13.14), to estimate the weights. Greene (1993) argues that one should not use the proportions pi ’s as estimates for the π i ’s because this is equivalent to using the yi2 ’s instead of σ 2i in the heteroskedastic regression. These will lead to inefficient estimates. If OLS on (13.14) is reported one should use the robust White heteroskedastic variance-covariance option, otherwise the standard errors are biased and inference is misleading. Ruhm (1996) uses a grouped logit analysis to study the impact of beer taxes and a variety of alcohol-control policies on motor vehicle fatality rates. Ruhm uses panel data of 48 states (excluding Alaska, Hawaii and the District of Columbia) over the period 1982-1988. The dependent variable is log[p/(1 − p)] where p is the total vehicle fatality rate per capita for state i at time t. The explanatory variables include the real beer tax rate on 24 (12 oz.) containers of beer, the minimum legal drinking age (MLDA) in years and five other dummy variables indicating the presence of alcohol regulations. These include BREATH test laws which is a dummy variable that takes the value 1 if the state authorized the police to administer pre-arrest breath


328

Chapter 13: Limited Dependent Variables

test to establish probable cause for driving under the influence (DUI). DRAMLAW is dummy variable taking the value 1 if the state has a statute or case law authorizing parties injured by an intoxicated driver to file a law suit against the alcohol server. PER SE takes the value 1 if the state licensing agency is required to suspend or revoke the driver’s license after arrest for DUI. CONSENT takes the value 1 if the state has a law requiring license sanction for refusing to submit to alcohol testing. JAIL which takes the value of 1 if the state passed legislation mandating jail or community service for the first DUI conviction. Other variables included are the unemployment rate, real per capita income and state and time dummy variables. This is the fixed effects specification discussed in Chapter 12. Details on the definitions of these variables are given in Table 1 of Ruhm (1996). Results showed that most of the regulations had little or no impact on traffic mortality. By contrast, higher beer taxes were associated with reductions in crash deaths. Table 13.1 shows the grouped logit regression results for the Ruhm (1996) data set using state and time dummy variables. The beer tax is negative and significant, while the minimum legal drinking age is not significant. Neither is the breath test law, the PER SE, CONSENT or JAIL variables, all of which represent state alcohol safety related legislation. However, DRAMLAW is negative and significant. Income per capita is positive and significant while the unemployment rate is negative and significant. Lower unemployment and higher income are signs of good economic conditions with the associated higher traffic density. Also, higher income per capita leads to higher consumption of alcohol, and higher unemployment leads to more depressed and unsatisfied labor force which could encourage more alcohol consumption and traffic related deaths. The state dummy variables are jointly significant with an observed F -value of 48.85 which is distributed as F (47, 272). The year dummies are jointly significant with an observed F -value of 8.01 which is distributed as F (6, 272). Problem 12 asks the reader to replicate Table 13.1. These results imply that increasing the minimum legal drinking age, revoking driver’s license or imposing stiffer punishments like mandating jail or community service are not effective policy tools for decreasing traffic related deaths. However, increasing the real tax on beer is an effective policy for reducing traffic related deaths. For grouped data, the sample sizes ni for each group have to be sufficiently large. Also, the pi ’s cannot be zero or one. One modification suggested in the literature is to add (1/2ni ) to pi when computing the log of odds ratio, see Cox (1970). Papke and Wooldridge (1996) argue that in many economic settings pi may be 0 or 1 for a large number of observations. For example, when studying participation rates in pension plans or when studying high school graduation rates. They propose a fractional logit regression which handles fractional response variables based on quasi-likelihood methods. Fractional response variables are bounded variables. Without loss of generality, they could be restricted to lie between 0 and 1. Examples include the proportion of income spent on charitable contributions, the fraction of total weekly hours spent working. Papke and Wooldridge (1996) propose modeling the E(yi /xi ) as a logistic function Λ(x′i β). This insures that the predicted value of yi lies in the interval (0, 1). It is also well defined even if yi takes the values 0 or 1 with positive probability. It is important to note that in case yi is a proportion from a group of known size ni , the quasi maximum likelihood method ignores the information on ni . Using the Bernoulli log-likelihood function, one gets Li (β) = yi log[Λ(x′i β)] + (1 − yi ) log[1 − Λ(x′i β)] for i = 1, 2, . . . , n, with 0 < Λ(x′i β) < 1.


13.4

Grouped Data

329

Table 13.1 Grouped Logit, Beer Tax and Motor Vehicle Fatality Fixed-effects (within) regression Group variable (i): state R-sq within = 0.3566 between = 0.2728 overall = 0.1735 corr(u i, Xb) = -0.7400 vfrall

Coef.

Std. Err.

t

P >t

beertax mlda breath dramlaw perse impcond jailcom incperc unrate year83 year84 year85 year86 year87 year88 cons

−.1609269 −.0008891 .0018155 −.0789037 −.0192373 .0232511 .0058019 .0246669 −.0335797 −.0336748 −.1028847 −.1348613 −.1005879 −.1337847 −.159083 −8.394118

.0782078 .0085417 .0226649 .0246141 .0199354 .0386605 .0282758 .0102202 .0050031 .0156339 .0194028 .0205257 .0224115 .0258383 .0292153 .2275537

−2.06 −0.10 0.08 −3.21 −0.96 0.60 0.21 2.41 −6.71 −2.15 −5.30 −6.57 −4.49 −5.18 −5.45 −36.89

0.041 0.917 0.936 0.002 0.335 0.548 0.838 0.016 0.000 0.032 0.000 0.000 0.000 0.000 0.000 0.000

sigma u sigma e rho

.36371961 .07263869 .9616454

F test that all u i=0:

Number of obs = Number of groups =

335 48

Obs per group: min = avg = max =

6 7.0 7

F(15,272)

=

10.05

Prob > F

=

0.0000

[95% Conf. Interval] −.3148965 −.0177053 −.0428055 −.1273621 −.0584845 −.0528608 −.0498654 .0045461 −.0434294 −.0644536 −.1410836 −.1752707 −.14471 −.1846531 −.2165999 −8.842108

−.0069573 .0159272 .0464365 −.0304453 .0200099 .099363 .0614693 .0447877 −.0237299 −.002896 −.0646859 −.0944519 −.0564658 −.0829162 −.1015661 −7.946127

(fraction of variance due to u i) F(47, 272) = 48.85

Prob > F = 0.0000

 √ Maximizing ni=1 Li (β) with respect to β yields the quasi-MLE which is consistent and n asymptotically normal regardless of the distribution of yi conditional on xi , see Gourieroux, Monfort and Trognon (1984) and McCullagh and Nelder (1989). The latter proposed the generalized linear models (GLM) approach to this problem in statistics. Logit QMLE can be done in Stata using the GLM command with the Binary family function indicating Bernoulli and the Link function indicating the logistic distribution. Papke and Wooldridge (1996) derive robust asymptotic variance of the QMLE of β and suggest some specification tests based on Wooldridge (1991). They apply their methods to the participation in 401(K) pension plans. The data are from the 1987 IRS Form 5500 reports of pension plans with more than 100 participants. This data set containing 4734 observations can be downloaded from the Journal of Applied Econometrics Data Archive. We focus on a subset of their data which includes 3874 observations of plans with match rates less than or equal to one. Match rates above one may be indicating end-of-plan year employer contributions made to avoid IRS disqualification. Participation rates (PRATE) in this sample are high


330

Chapter 13: Limited Dependent Variables

Table 13.2 Logit Quasi-MLE of Participation Rates in 401(K) Plan glm prate mrate log emp log emp2 age age2 sole if one==1, f(bin) l(logit) robust note: prate has non-integer values Iteration Iteration Iteration Iteration

0: 1: 2: 3:

log log log log

pseudo-likelihood pseudo-likelihood pseudo-likelihood pseudo-likelihood

= = = =

Generalized linear models Optimization : ML: Newton-Raphson Deviance Pearson

= 1273.60684 = 724.4199889

Variance function Link function Standard errors

: V(u) = u*(1-u) : g(u) = ln(u/(1-u)) : Sandwich

−1200.8698 −1179.3843 −1179.2785 −1179.2785

mrate log emp log emp2 age age2 sole cons

= 3784 = 3777 = 1 = .3372006 = .1917977

[Bernoulli] [Logit]

Log pseudo-likelihood = −1179.278516 BIC = −29843.34715 prate

Number of obs Residual df Scale parameter (1/df) Deviance (1/df) Pearson

AIC

Coef.

Robust Std. Err.

z

P > |z|

1.39008 −1.001874 .0521864 .0501126 −.0005154 .0079469 5.057997

.1077064 .1104365 .0071278 .0088451 .0002117 .0502025 .4208646

12.91 −9.07 7.32 5.67 −2.43 0.16 12.02

0.000 0.000 0.000 0.000 0.015 0.874 0.000

= .6269971 [95% Conf. Interval] 1.17898 −1.218326 .0382161 .0327766 −.0009303 −.0904482 4.233117

1.601181 −.7854229 .0661568 .0674486 −.0001004 .1063421 5.882876

averaging 84.8%. Over 40% of the plans have a participation proportion of one. This makes the log-odds ratio approach awkward since adjustments have to be made to more than 40% of the observations. The plan match rate (MRATE) averages about 41 cents on the dollar. Other explanatory variables include total firm employment (EMP), age of the plan (AGE), a dummy variable (SOLE) which takes the value of 1 if the 401(K) plan is the only pension plan offered by the employer. The 401(K) plans average 12 years in age, they are the SOLE plan in 37% of the sample. The average employment is 4622. Problem 14 asks the reader to replicate the descriptive statistic given in Table I of Papke and Wooldridge (1996, p. 627). Table 13.2 gives the Stata output for logit QMLE using the same specification given in Table II of Papke and Wooldridge (1996, p. 628). Note that it uses the GLM command, the Bernoulli variance function and the logit link function. The results show that there is a positive and significant relationship between match rate and participation rate. All the other variables included are significant except for SOLE. Problem 14 asks the reader to replicate this result and compare with OLS. The latter turns out to have a lower R2 and fails a RESET test, see Chapter 8.


13.5

13.5

Individual Data: Probit and Logit

331

Individual Data: Probit and Logit

When the number of observations ni in each group is small, one cannot obtain reliable estimates of the π i ’s with the pi ’s. In this case, one should not group the observations, instead these observations should be treated as individual observations and the model estimated by the maximum likelihood procedure. The likelihood is obtained as independent random draws from a Bernoulli distribution with probability of success π i = F (x′i β) = P [yi = 1]. Hence ℓ=

n

′ yi i=1 [F (xi β)] [1

− F (x′i β)]1−yi

(13.15)

and the log-likelihood logℓ =

n

′ i=1 {yi logF (xi β)

+ (1 − yi )log[1 − F (x′i β)]}

(13.16)

The first-order conditions for maximization require the score S(β) = ∂logℓ/∂β to be zero:  S(β) = ∂logℓ/∂β = ni=1 {[fi yi /Fi ] − (1 − yi )[fi /(1 − Fi )]}xi (13.17) n = i=1 (yi − Fi )fi xi /[Fi (1 − Fi )] = 0

where the subscript i on f or F denotes that the argument of that function is x′i β . For the logit model (13.17) reduces to S(β) =

n

i=1 (yi

− Λi )xi = 0

since

fi = Λi (1 − Λi )

(13.18)

 If there is a constant in the model, the solution to (13.18) for xi = 1 implies that ni=1 yi = n  i=1 Λi . This means that the number of participants in the sample, i.e., those with yi = 1, will always be equal to the predicted number of participants from the logit model. Similarly, if xi is a dummy variable which is 1 if the individual is male and zero if the individual is female, then (13.18) states that the predicted frequency is equal to the actual frequency for males and  i ) as females. Note that (13.18) resembles the OLS normal equations if we interpret (yi − Λ residuals. For the probit model (13.17) reduces to n S(β) = (yi − Φi )φi xi /[Φi (1 − Φi )] (13.19) i=1  λ x + λ x = 0 = yi =0 oi i yi =1 1i i

where λoi = −φi /[1−Φi ] for yi = 0 and λ1i = φi /Φi for yi = 1. Also, Σyi =0 denotes the sum over all zero values of yi . These λi ’s are thought of as generalized residuals which are orthogonal to xi . Note that unlike the logit, the probit does not necessarily predict the number of participants to be exactly equal to the number of ones in the sample. Equations (13.17) are highly nonlinear and may be solved using the scoring method, i.e., starting with some initial value β o we revise this estimate as follows: β 1 = β o + [I −1 (β o )]S(β o )

(13.20)

where S(β) = ∂logℓ/∂β and I(β) = E[−∂ 2 logℓ/∂β∂β ′ ]. This process is repeated until convergence. For the logit and probit models, logF (x′i β) and log[1 − F (x′i β)] are concave. Hence, the log-likelihood function given by (13.16) is globally concave, see Pratt (1981). Hence, for both the logit and probit, [∂ 2 logℓ/∂β∂β ′ ] is negative definite for all values of β and the iterative  procedure will converge to the unique maximum likelihood estimate β M LE no matter what


332

Chapter 13: Limited Dependent Variables

 M LE is estimated by starting values we use. In this case, the asymptotic covariance matrix of β −1  I (β M LE ) from the last iteration. Amemiya (1981, p. 1495) derived I(β) by differentiating (13.17), multiplying by a negative sign and taking the expected value, the result is given by:  (13.21) I(β) = −E[∂ 2 logℓ/∂β∂β ′ ] = ni=1 fi2 xi x′i /Fi (1 − Fi ) For the logit, (13.21) reduces to  I(β) = ni=1 Λi (1 − Λi )xi x′i

For the probit, (13.21) reduces to  I(β) = ni=1 φ2i xi x′i /Φi (1 − Φi )

(13.22)

(13.23)

Alternative maximization may use the Newton-Raphson iterative procedure which uses the Hessian itself rather than its expected value in (13.20), i.e., I(β) is replaced by H(β) = [−∂ 2 logℓ/∂β∂β ′ ].For the logit model, H(β) = I(β) and is given in (13.22). For the probit n 2 ′ ′ model, H(β) = i=1 [λi + λi xi β]xi xi which is different from (13.23). Note that λi = λoi if yi = 0; and λi = λ1i if yi = 1. These were defined below (13.19). A third method, suggested by Berndt, Hall, Hall and Hausman (1974) uses the outer product ′ (β). For the logit model, this is of the first of I(β), i.e., G(β) = S(β)S n derivatives2 in place ′ G(β) = i=1 (yi − Λi ) xi xi . For the probit model, G(β) = ni=1 λ2i xi x′i . As in the method of scoring, one iterates starting from initial estimates β o , and the asymptotic variance-covariance  H(β)  or I(β)  in the last iteration. matrix is estimated from the inverse of G(β), Test of hypotheses can be carried out from the asymptotic standard errors using t-statistics.  − r)′ [RV (β)R  ′ ]−1 (Rβ  − r) can For Rβ = r type restrictions, the usual Wald test W = (Rβ  obtained from the last iteration as described above. Likelihood ratio and be used with V (β) Lagrange Multiplier statistics can also be computed. LR = −2[logℓrestricted − logℓunrestricted ], whereas, the Lagrange Multiplier statistic is LM = S ′ (β)V (β)S(β), where S(β) is the score  based evaluated at the restricted estimator. Davidson and MacKinnon (1984) suggest that V (β)  on I(β) is the best of the three estimators to use. In fact, Monte Carlo experiments show that  based on the outer product of the first derivatives usually performs the the estimate of V (β) worst and is not recommended in practice. All three statistics are asymptotically equivalent and are asymptotically distributed as χ2q where q is the number of restrictions. The next section discusses tests of hypotheses using an artificial regression.

13.6

The Binary Response Model Regression1

Davidson and MacKinnon (1984) suggest a modified version of the Gauss-Newton regression (GNR) considered in Chapter 8 which is useful in the context of a binary response model described in (13.5).2 In fact, we have shown that this model can be written as a nonlinear regression yi = F (x′i β) + ui

(13.24)

with ui having zero mean and var(ui ) = Fi (1 − Fi ). The GNR ignoring heteroskedasticity yields (yi − Fi ) = fi x′i b + residual


13.6

The Binary Response Model Regression

333

where b is the regression estimates when we regress (yi − Fi ) on fi x′i . Correcting for heteroskedasticity by dividing each observation by its standard deviation we get the Binary Response Model Regression (BRMR): fi (y − Fi )  i = x′i b + residual Fi (1 − Fi ) Fi (1 − Fi )

(13.25)

For the logit model with fi = Λi (1 − Λi ), this simplifies further to yi − Λi  ′ √ = fi xi b + residual fi

(13.26)

For the probit model, the BRMR is given by 

φi yi − Φi = x′i b + residual Φi (1 − Φi ) Φi (1 − Φi )

(13.27)

Like the GNR considered in Chapter 8, the BRMR given in (13.25) can be used for obtaining parameter and covariance matrix estimates as well as test of hypotheses. In fact, Davidson and MacKinnon point out that the transpose of the dependent variable in (13.25) times the matrix of regressors in (13.25) yields a vector whose typical element is exactly that of S(β) given in (13.17). Also, the transpose of the matrix of regressors in (13.25) multiplied by itself yields a matrix whose typical element is exactly that of I(β) given in (13.21). Let us consider how the BRMR is used to test hypotheses. Suppose that β ′ = (β ′1 , β ′2 ) where ′ = (β  , 0) β 1 is of dimension k − r and β 2 is of dimension r. We want to test Ho ; β 2 = 0. Let β 1 be the restricted MLE of β subject to Ho . In order to test Ho , we run the BRMR: 

yi − Fi

Fi (1 − Fi )

=

fi

Fi (1 − Fi )

x′i1 b1 + 

fi

Fi (1 − Fi )

x′i2 b2 + residual

(13.28)

where x′i = (x′i1 , x′i2 ) has been partitioned into vectors conformable with the corresponding  and fi = f (x′ β).  The suggested test statistic for Ho is the partition of β. Also, Fi = F (x′i β) i explained sum of squares of the regression (13.28). This is asymptotically distributed as χ2r under Ho .3 A special case of this BRMR is that of testing the null hypothesis that all the slope coefficients are zero. In this case, xi1 = 1 and β 1 is the constant α. Problem 2 shows that the restricted MLE in this case is F (α) = y¯ or α  = F −1 (¯ y ), where y¯ is the proportion of the sample with yi = 1. Therefore, the BRMR in (13.25) reduces to 

fi ( α) fi ( α) yi − y¯ = b1 +  x′i2 b2 + residual y¯(1 − y¯) y¯(1 − y¯) y¯(1 − y¯)

(13.29)

Note that y¯(1 − y¯) is constant for all observations. The test for b2 = 0 is not affected by dividing the dependent variable or the regressors by this constant, nor is it affected by subtracting a constant from the dependent variable. Hence, the test for b2 = 0 can be carried out by regressing yi on a constant and xi2 and testing that the slope coefficients of xi2 are zero using the usual least squares F -statistic. This is a simpler alternative to the likelihood ratio test proposed in the previous section and described in the empirical example in section 13.9. For other uses of the BRMR, see Davidson and MacKinnon (1993).


334

Chapter 13: Limited Dependent Variables

13.7

Asymptotic Variances for Predictions and Marginal Effects

 and the marginal Two results of interest after estimating the model are: the predictions F (x′ β) ′   effects ∂F/∂x = f (x β)β. For example, given the characteristics of an individual x, we can predict his or her probability of purchasing a car. Also, given a change in x, say income, one can estimate the marginal effect this will have on the probability of purchasing a car. The latter effect is constant for the linear probability model and is given by the regression coefficient of income, whereas for the probit and logit models this marginal effect will vary with the xi ’s, see (13.7) and (13.8). These marginal effects can be computed with Stata using the dprobit command. The default is to compute them at the sample mean x. There is also the additional problem of computing variances for these predictions and marginal effects. Both  and f (x′ β)  β  are nonlinear functions of the β’s.  To compute standard errors, we can use F (x′ β)  then the asy.var( the following linear approximation which states that whenever  θ = F (β) θ) = ′ ′     and f =  (∂F/∂ β) V (β)(∂F/∂ β). For the predictions, let z = x β and denote by F = F (x′ β)  then f (x′ β),  = (∂ F /∂z)(∂z/∂ β)  = fx ∂ F /∂ β

and

 then For the marginal effects, let γ  = fβ, ′

 asy.var(F) = f2 x′ V (β)x.

 )V (β)(∂  γ /∂ β  )′ asy.var( γ ) = (∂ γ /∂ β

 f/∂z)(∂z/∂ β  ′ ) = fIk + (∂ f/∂z)(βx  ′ ).  ′ = fIk + β(∂ where ∂ γ /∂ β ′   So, ∂  = φ[I  k − z βx  ′ ] and For the probit model, ∂ f/∂z = ∂ φ/∂z = −z φ. γ /∂ β 2 [Ik − x′ β βx  ′ ]V (β)[I  k − x′ β  βx  ′ ]′ asy.var( γ) = φ

(13.30)

(13.31)

 − Λ),  so For the logit model, f = Λ(1

 Λ/∂z)   f) = (1 − 2Λ)  Λ(1  − Λ)  ∂ f/∂z = (1 − 2Λ)(∂ = (1 − 2Λ)( ′  ′]  = Λ(1  − Λ)[I  k + (1 − 2Λ)  βx ∂ γ /∂ β

and (13.30) becomes

 ′ ]V (β)[I  k + (1 − 2Λ)  ′ ]′  βx  βx  − Λ)]  2 [Ik + (1 − 2Λ) asy.var( γ ) = [Λ(1

13.8

(13.32)

Goodness of Fit Measures

There are problems with the use of conventional R2 -type measures when the explained variable y takes on two values, see Maddala (1983, pp. 37-41). The predicted values y are probabilities and the actual values of y are either 0 or 1 so the usual R2 is likely to be very if there low. Also, is a constant in the model the linear probability and logit models satisfy ni=1 yi = ni=1 yi . However, the probit model does not necessarily satisfy this exact relationship although it is approximately valid. Several R2 -type measures have been suggested in the literature, some of these are the following: 2 . (i) The squared correlation between y and y: R12 = ry, y


13.9

Empirical Examples

335

(ii) Measures based on the residual sum of squares: Effron (1978) suggested using    R22 = 1 − [ ni=1 (yi − yi )2 / ni=1 (yi − y¯)2 ] = 1 − [n ni=1 (yi − yi )2 /n1 n2 ]    y 2 = n1 − n(n1 /n)2 = n1 n2 /n, where n1 = ni=1 yi and since ni=1 (yi − y¯)2 = ni=1 yi2 − n¯ n2 = n − n1 .

n Amemiya (1981, p. 1504) suggests using i )2 / yi (1 − yi ) as the residual sum i=1 (yi − y of squares. This weights each squared error by the inverse of its variance.

(iii) Measures based on likelihood ratios: R32 = 1 − (ℓr /ℓu )2/n where ℓr is the restricted likelihood and ℓu is the unrestricted likelihood. This tests that all the slope coefficients are zero in the standard linear regression model. For the limited dependent variable model however, the likelihood function has a maximum of 1. This means that ℓr ≤ ℓu ≤ 1 or ℓr ≤ (ℓr /ℓu ) ≤ 2/n 2/n 1 or ℓr ≤ 1 − R32 ≤ 1 or 0 ≤ R32 ≤ 1 − ℓr . Hence, Cragg and Uhler (1970) suggest a 2/n 2/n 2/n 2/n pseudo-R2 that lies between 0 and 1, and is given by R42 = (ℓu − ℓr )/[(1 − ℓr )/ℓu ]. Another measure suggested by McFadden (1974) is R52 = 1 − (logℓu /logℓr ). (iv) Proportion of correct predictions: After computing y, one classifies the i-th observation as a success if yi > 0.5, and a failure if yi < 0.5. This measure is useful but may not have enough discriminatory power.

13.9

Empirical Examples

Example 1: Union Participation To illustrate the logit and probit models, we consider the PSID data for 1982 used in Chapter 4. In this example, we are interested in modelling union participation. Out of the 595 individuals observed in 1982, 218 individuals had their wage set by a union and 377 did not. The explanatory variables used are: years of education (ED), weeks worked (WKS), years of full-time work experience (EXP), occupation (OCC = 1, if the individual is in a blue-collar occupation), residence (SOU T H = 1, SM SA = 1, if the individual resides in the South, or in a standard metropolitan statistical area), industry (IN D = 1, if the individual works in a manufacturing industry), marital status (M S = 1, if the individual is married), sex and race (F EM = 1, BLK = 1, if the individual is female or black). A full description of the data is given in Cornwell and Rupert (1988). The results of the linear probability, logit and probit models are given in Table 13.3. These were computed using EViews. In fact Table 13.4 gives the probit output. We have already mentioned that the probit model normalizes σ to be 1. But, the logit model has variance π 2 /3. Therefore, the√logit estimates tend to be larger than the probit estimates although by a factor less than π/ 3. In order to make the logit results comparable to those of the probit, Amemiya (1981) suggests multiplying the logit coefficient estimates by 0.625. Similarly, to make the linear probability estimates comparable to those of the probit model one needs to multiply these coefficients by 2.5 and then subtract 1.25 from the constant term. For this example, both logit and probit procedures converged quickly in 4 iterations. The loglikelihood values and McFadden’s (1974) R2 obtained for the last iteration are recorded. Note that the logit and probit estimates yield similar results in magnitude, sign and significance. One would expect different results from the logit and probit only if there are several observations in the tails. The following variables were insignificant at the 5% level: EXP, IND,


336

Chapter 13: Limited Dependent Variables

Table 13.3 Comparison of the Linear Probability, Logit and Probit Models: Union Participation∗ Variable EXP WKS OCC IND SOUTH SMSA MS FEM ED BLK Const Log-likelihood McFadden’s R2 χ210 ∗

OLS

Logit

Probit

−.005 (1.14) −.045 (5.21) .795 (6.85) .075 (0.79) −.425 (4.27) .211 (2.20) .247 (1.55) −.272 (1.37) −.040 (1.88) .125 (0.71) 1.740 (5.27)

−.007 (1.15) −.068 (5.05) 1.036 (6.27) .114 (0.89) −.653 (4.33) .280 (2.05) .378 (1.66) −.483 (1.58) −.057 (1.85) .222 (0.90) 2.738 (3.27) −312.337 0.201 157.2

−.007 (1.21) −.061 (5.16) .955 (6.28) .093 (0.76) −.593 (4.26) .261 (2.03) .351 (1.62) −.407 (1.47) −.057 (1.99) .226 (0.99) 2.517 (3.30) −313.380 0.198 155.1

Figures in parentheses are t-statistics

MS, FEM and BLK. The results show that union participation is less likely if the individual resides in the South and more likely if he or she resides in a standard metropolitan statistical area. Union participation is also less likely the more the weeks worked and the higher the years of education. Union participation is more likely for blue-collar than non blue-collar occupations. The linear probability model yields different estimates from the logit and probit results. OLS predicts two observations with yi > 1, and 29 observations with yi < 0. Table 13.5 gives the actual versus predicted values of union participation for the linear probability, logit and probit models. The percentage of correct predictions is 75% for the linear probability and probit model and 76% for the logit model. One can test the significance of all slope coefficients by computing the LR based on the unrestricted log-likelihood value (logℓu ) reported in Table 13.3, and the restricted log-likelihood value including only the constant. The latter is the same for both the logit and probit models and is given by y log¯ y + (1 − y¯)log(1 − y¯)] logℓr = n[¯

(13.33)

where y¯ is the proportion of the sample with yi = 1, see problem 2. In this example, y¯ = 218/595 = 0.366 and n = 595 with logℓr = −390.918. Therefore, for the probit model, LR = −2[logℓr − logℓu ] = −2[−390.918 + 313.380] = 155.1

which is distributed as χ210 under the null of zero slope coefficients. This is highly significant and the null is rejected. Similarly, for the logit model this LR statistic is 157.2. For the linear probability model, the same null hypothesis of zero slope coefficients can be tested using a Chow F -statistic. This yields an observed value of 17.80 which is distributed as F (10, 584) under the null hypothesis. Again, the null is soundly rejected. This F -test is in fact the BRMR test considered in section 13.6. As described in section 13.8, McFadden’s R2 is given by R52 = 1 − [logℓu /logℓr ] which for the probit model yields R52 = 1 − (313.380/390.918) = 0.198.

For the logit model, McFadden’s R52 is 0.201.


13.9

Empirical Examples

337

Table 13.4 Probit Estimates: Union Participation Dependent Variable: UNION Method: ML – Binary Probit Sample: 1 595 Included observations: 595 Convergence achieved after 5 iterations Covariance matrix computed using second derivatives Variable

Coefficient

Std. Error

z-Statistic

Prob.

EX WKS OCC IND SOUTH SMSA MS FEM ED BLK C

–0.006932 –0.060829 0.955490 0.092827 –0.592739 0.260700 0.350520 –0.407026 –0.057382 0.226482 2.516784

0.005745 0.011785 0.152137 0.122774 0.139102 0.128630 0.216284 0.277038 0.028842 0.228845 0.762612

–1.206491 –5.161666 6.280476 0.756085 –4.261183 2.026741 1.620648 –1.469203 –1.989515 0.989675 3.300217

0.2276 0.0000 0.0000 0.4496 0.0000 0.0427 0.1051 0.1418 0.0466 0.3223 0.0010

Mean dependent var S.E. of regression Sum squared resid Log likelihood Restr. log likelihood LR statistic (10 df) Probability(LR stat)

0.366387 0.420828 103.4242 –313.3795 –390.9177 155.0763 0.000000

S.D. dependent var Akaike info criterion Schwarz criterion Hannan-Quinn criter. Avg. log likelihood McFadden R-squared

Obs with Dep=0 Obs with Dep=1

377 218

Total obs

0.482222 1.090351 1.171484 1.121947 –0.526688 0.198349

595

Table 13.5 Actual versus Predicted: Union Participation Predicted Union = 0

Total

Union = 1

Union =0

OLS = LOGIT = Probit =

312 316 314

OLS = LOGIT = Probit =

65 61 63

377

Union =1

OLS = LOGIT = Probit =

83 82 86

OLS = LOGIT = Probit =

135 136 132

218

OLS = LOGIT = Probit =

395 398 400

OLS = LOGIT = Probit =

200 197 195

595

Actual

Total


338

Chapter 13: Limited Dependent Variables

Table 13.6 Probit Estimates: Employment and Problem Drinking Dependent Variable: EMPL Method: ML – Binary Probit Sample: 1 9822 Included observations: 9822 Convergence achieved after 7 iterations QML (Huber/White) standard errors & covariance Variable

Coefficient

Std. Error

z-Statistic

Prob.

90th Pctl. UE88 AGE AGE2 SCHOOLING MARRIED FAMILY SIZE WHITE EXCELLENT VERY GOOD GOOD FAIR NORTH EAST MIDWEST SOUTH CENTER OTHER MSA QU1 QU2 QU3 CONSTANT

−0.104947 −0.053277 0.099634 −0.001304 0.047183 0.295292 0.018891 0.394523 1.816306 1.778434 1.547836 1.043363 0.034312 0.060491 0.182121 −0.073053 0.075953 −0.105484 −0.051323 −0.029342 −3.017454

0.058985 0.014202 0.017118 0.000205 0.006674 0.054083 0.014046 0.048336 0.09834 0.099148 0.098259 0.107722 0.061999 0.053786 0.054232 0.051869 0.051307 0.05277 0.052816 0.054372 0.359214

−1.779203 −3.751467 5.820531 −6.359957 7.070238 5.45998 1.344943 8.162155 18.46972 17.93715 15.75266 9.685669 0.553434 1.124662 3.358186 −1.408406 1.480381 −1.998942 −0.971734 −0.539648 −8.400162

0.0752 0.0002 0.0000 0.0000 0.0000 0.0000 0.1786 0.0000 0.0000 0.0000 0.0000 0.0000 0.5800 0.2607 0.0008 0.1590 0.1388 0.0456 0.3312 0.5894 0.0000

Mean dependent var S.E. of regression Sum squared resid Log likelihood Avg. log likelihood

0.898188 0.277414 754.2706 −2698.180 −0.274708

Obs with Dep=0 Obs with Dep=1

1000 8822

S.D. dependent var Akaike info criterion Schwarz criterion Hannan-Quinn criter. Total obs

0.302417 0.553692 0.569069 0.558902 9822

Example 2: Employment and Problem Drinking Mullahy and Sindelar (1996) estimate a linear probability model relating employment and measures of problem drinking. The analysis is based on the 1988 Alcohol Supplement of the National Health Interview Survey. This regression was performed for Males and Females separately since the authors argue that women are less likely than men to be alcoholic, are more likely to abstain from consumption, and have lower mean alcohol consumption levels. They also report that women metabolize ethanol faster than do men and experience greater liver damage for the same level of consumption of ethanol. The dependent variable takes the value 1 if the individual was employed in the past two weeks and zero otherwise. The explanatory variables included


13.10

Multinomial Choice Models

339

the 90th percentile of ethanol consumption in the sample (18 oz. for males and 10.8 oz. for females) and zero otherwise. The state unemployment rate in 1988 (UE88), Age, Age2 , schooling, married, family size, and white. Health status dummies indicating whether the individual’s health was excellent, very good, fair. Region of residence, whether the individual resided in the northeast, midwest or south. Also, whether he or she resided in center city or other metropolitan statistical area (not center city). Three additional dummy variables were included for the quarters in which the survey was conducted. Details on the definitions of these variables are given in Table 1 of Mullahy and Sindelar (1996). Table 13.6 gives the probit results based on n = 9822 males. These results show a negative relationship between the 90th percentile alcohol variable and the probability of being employed, but this has a p-value of 0.075. Mullahy and Sindelar find that for both men and women, problem drinking results in reduced employment and increased unemployment. Problem 13 asks the reader to verify these results as well as those in the original article by Mullahy and Sindelar (1996).

13.10

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